I Chronic Diseases in Canada n this is sue

I Chronic  Diseases in  Canada n this is sue
Chronic Diseases
in Canada
Volume 20, No 2
1999
In this issue
51
Prevalence and Geographic Disparities in Certain Congenital Anomalies in
Quebec: Comparison of Estimation Methods
Robert Choinière, Michel Pageau and Marc Ferland
58
Monograph Series on Aging-related Diseases: XII. Parkinson’s
Disease—Recent Developments and New Directions
Natalie Kontakos and Julie Stokes
77
Development of Record Linkage of Hospital Discharge Data for the Study of
Neonatal Readmission
Shiliang Liu and Shi Wu Wen
82
Rate and Cost of Hospitalizations for Asthma in Quebec: An Analysis of
1988/89, 1989/90 and 1994/95 Data
Claudine Laurier, Wendy Kennedy, Jean-Luc Malo, Michèle Paré, Daniel Labbé, André Archambault
and André-Pierre Contandriopoulos
89
The Cost of Suicide Mortality in New Brunswick, 1996
96
Workshop Report
Dale Clayton and Alberto Barceló
Canadian National Workshop on Measurement of Sun-related Behaviours
Chris Lovato, Jean Shoveller, Christina Mills and an Expert Panel
(continued on reverse)
Our mission is to help the people of Canada
maintain and improve their health.
Health Canada
(Contents continued)
101
Book Review
Injury Prevention: An International Perspective
Epidemiology, Surveillance, and Policy
Reviewed by Margaret Herbert
102
Calendar of Events
5
Information for Authors (on inside back cover)
Published by authority of the Minister of Health
© Minister of Public Works and Government Services Canada 1999
ISSN 0228-8699
Aussi disponible en français sous le titre Maladies chroniques au Canada
Prevalence and Geographic Disparities in Certain
Congenital Anomalies in Quebec: Comparison of
Estimation Methods
Robert Choinière, Michel Pageau and Marc Ferland
Abstract
The purpose of this study was to estimate the prevalence of congenital anomalies in Quebec
from MED-ECHO hospitalization records and from records of stillbirths. The results are first
compared with those from the Canadian Congenital Anomalies Surveillance System (CCASS)
for Quebec and Canada; then the data are examined by period and region of residence. The
study results show that, for the congenital anomalies selected for the study, the prevalence
rates measured for Quebec from the MED-ECHO data tend to be lower than the prevalence
rates for Canada, whereas the rates estimated by CCASS are higher for Quebec than for
Canada. The MED-ECHO data cover practically all Quebec births, compared with only 15%
coverage by CCASS, and therefore provide a more accurate picture of congenital anomalies
in Quebec.
Key words: congenital anomalies; disparities; estimates; evolution; prevalence; Quebec
Background
In its 1992 health and welfare policy, Quebec set
itself the goal of reducing the incidence of congenital
anomalies. In Quebec, congenital anomalies are the
second leading cause of perinatal death and the sixth
leading cause, in terms of years of potential life lost, of
premature death.1 Approximately 40% of all babies who
die during the first year of life and over 30% of children
admitted to a hospital pediatric department have a
congenital defect.1 Furthermore, it has been shown that
children born with congenital anomalies are much more
likely than others to suffer adverse long-term
consequences to their health, quality of life and survival,
such as prolonged periods in hospital; multiple surgeries;
disrupted physical, intellectual or motor development;
and respiratory, visual, auditory or language disorders.1–3
Since there is no registration or surveillance system
for congenital anomalies in Quebec, there is very little
accurate, recent information on the overall incidence of
the births of children with congenital defects.1,4 Thus, it
is difficult to know whether Quebec’s goal of reducing
congenital anomalies is being achieved. In Canada, the
only source of information on the birth prevalence of
congenital anomalies is the Canadian Congenital
Anomalies Surveillance System (CCASS), which for
several years has been administered through the
Laboratory Centre for Disease Control (LCDC), Health
Canada.5 However, the data are not fully representative
of Quebec.
CCASS uses provincial data on cases of congenital
defects obtained exclusively from hospital
admission/separation records of stillborns, newborns and
infants during the first year of life.6 CCASS does not
include cases associated with medical termination of
pregnancy for congenital defects, or with miscarriages
and abortions, and this limits the coverage of the
prevalence of congenital anomalies.
Since there may be several hospitalizations per
individual, CCASS uses a melding process to combine
information compiled on a single patient during different
hospital stays. This process matches variables such as
sex, date of birth, residential postal code and health
insurance number.
The portion of Quebec hospitalization records
compiled by MED-ECHO (the Quebec hospitalization
Author References
Robert Choinière, Régie régionale de la santé et des services sociaux (RRSSS) de Montréal-Centre, Direction de la santé publique, 3725, rue
Saint-Denis, Montréal (Québec) H2X 3L9; Fax: (514) 286-5782; E-mail: Robert_Choiniè[email protected]
Michel Pageau and Marc Ferland, RRSSS de Québec, Direction de la santé publique
1999
51
database) and sent to Health Canada does not contain all
the information needed to identify individuals and meld
cases.5 Therefore, LCDC incorporates into CCASS the
Quebec hospitalization data from the Hospital Medical
Records Institute (HMRI), an organization that for
several years has been administered by the Canadian
Institute for Health Information. These data have the
advantage of containing the information needed for
melding, but the major disadvantage of covering only a
small proportion of hospital births in Quebec. From 1989
to 1991, only 15% of Quebec births were captured by the
HMRI database,5 whereas from 1989 to 1995, 99% of
live births were registered by MED-ECHO. The CCASS
data for Quebec are therefore incomplete and cannot be
used to estimate the number of congenital anomalies or
accurately measure prevalence.
The goal of this study was to estimate the prevalence
of congenital anomalies in Quebec using MED-ECHO
data, which are more complete than the HMRI data used
by CCASS. We had access to several variables that are
not included in the version of MED-ECHO data sent to
Health Canada. This enabled us to use identifiers
comparable to the ones available from the HMRI
database while covering almost all births in Quebec. By
combining these data with those from the records of
stillbirths, we therefore expect to obtain more accurate
results and statistically more robust rates, because they
are based on a larger number of events. A deterministic
melding method was selected, which is different from,
but comparable to, the probabilistic one used in CCASS.
Methods
Data
To estimate the prevalence of congenital anomalies in
Quebec as accurately as possible, we used the congenital
defects cases from the MED-ECHO hospitalization
database and from stillbirth records, which MED-ECHO
does not cover. MED-ECHO is the Quebec database on
short-term hospitalizations and day surgeries performed
in Quebec. Each record contains identifying,
demographic information along with the primary
diagnosis on admission and 15 possible secondary
diagnoses.4,7 As in the case of CCASS, it was not
possible to include cases associated with medical
terminations of pregnancy for congenital defects, or with
miscarriages and abortions.
The data relating to an individual can be melded using
the probabilistic or the deterministic method, depending
on the quantity and quality of the information available
for correctly identifying individuals.8 When little
information is available, as is the case for CCASS, a
probabilistic approach is recommended (based, as its
name suggests, on probabilities). To build its database,
CCASS uses a variation of this method that involves ad
hoc weighting based on a system of weighting factors.9
52
Chronic Diseases in Canada
When a large amount of high-quality data is available,
a deterministic approach that makes links according to
criteria established by the researchers is often preferable.
The deterministic method has the advantage of being
easier to apply while producing better results than the
probabilistic approach.10 For these reasons, we chose the
deterministic approach for this study.
The variables that CCASS selects from MED-ECHO
to meld hospitalizations are sex, date of birth, six-digit
postal code, health insurance number and the codes from
the Ninth Revision of the International Classification of
Diseases (ICD-9) for the main diagnosis and the 15
secondary diagnoses. To maximize melds, the following
variables were added to these identifiers: admission date,
discharge date, medical record number, hospital code,
municipality code, regional county municipality code
(MRC), local community services centre code (CLSC),
type of death, type of care, civil status, origin code and
destination code.
Using a deterministic melding procedure, we
estimated the number of infants who were hospitalized
for congenital anomalies at least once during their first
year of life by regrouping the different hospitalization
records for each child under the age of one. This was
done for the fiscal years from 1988/89 to 1996/97.
The first step consisted of matching hospitalizations
by file number, because only one file number can be
assigned to an individual in the same hospital no matter
how many visits are made. In the second step,
hospitalizations with identical health insurance numbers
were matched, as these represent another identifier
unique to every individual. However, health insurance
numbers are frequently available only several months
after the child’s birth and therefore cannot be used to
link visits occurring in the first few months of life.
Finally, in the later stages, the following four remaining
matches were made based on the place of residence of
the hospitalized individuals.
• Identical postal code (six digits), date of birth and sex
• Identical municipality code, CLSC code, date of birth
and sex
• Identical postal code (three digits), CLSC code, date
of birth and sex
• Identical postal code (three digits), municipality code,
date of birth and sex
For the last four melding procedures to be acceptable,
each match had to meet the following conditions: the
civil status for a second visit could not be “newborn,” the
discharge from a first visit could not be due to death, the
admission date for a second visit could not precede by
more than one day the discharge date of the preceding
visit and, if the admissions were to the same hospital, the
file numbers could not be different.
Vol 20, No 2
TABLE 1
TABLE 2
Melding procedures for 79,409 hospitalizations for
congenital anomalies and other ICD-9 codes
selected by CCASS, 1989–1995a
Codes from the Ninth Revision of the
International Classification of
Diseases (ICD-9) for selected
congenital anomalies
Procedure
Hospitalizations
eliminated
Number of
hospitalizations
after each step
MELD I
8,228 duplicates eliminated
From the file number and
hospital
71,181
MELD II
From the health
insurance number
470 duplicates eliminated
70,711
MELD III
From the postal code (6
digits), date of birth and
sex
4,448 duplicates eliminated
66,263
MELD IV
1,744 duplicates eliminated
From the CLSC,
municipality, date of birth
and sex
64,519
MELD V
343 duplicates eliminated
From the postal code (3
digits), the CLSC, date of
birth and sex
64,176
MELD VI
45 duplicates eliminated
From the postal code (3
digits), municipality, date
of birth and sex
64,131
FINAL STEP
4,975 hospitalizations with
no diagnosis of congenital
Deletion of information
anomalies
unrelated to congenital
anomalies and correction
of contradictory
diagnoses
59,156
(hospitalizations
for congenital
anomalies)
For matches based on the file number and the health
insurance number where the date of birth varied between
the two visits, we assumed that the date on the first visit
was the most accurate. When the sex varied from one
visit to another, we used the sex from the most recent
visit.
The final operation needed to create the congenital
abnormalities database for Quebec was to remove certain
cases of contradictory diagnoses using information
previously provided by CCASS. Some cases initially
identified as congenital anomalies were also associated
with particular diagnoses that cancelled the initial
diagnosis. Finding one of these diagnoses in addition to a
diagnosis of congenital anomalies for the same
hospitalization implied there was a contradiction among
diagnoses and that this hospitalization, according to
CCASS, should not be included among the congenital
anomalies.
1999
Congenital anomaly
ICD-9 code
Anencephalus and similar
anomalies
740.0–740.2
Spina bifida
741.0–741.9
Encephalocele
742.0
Congenital hydrocephalus
742.3
Transposition of great vessels
745.1
Hypoplastic left heart syndrome
746.7
Cleft palate
749.0
Cleft palate with cleft lip
749.2
Tracheo-esophageal fistula,
esophageal atresia and stenosis
750.3
Atresia and stenosis of large
intestine, rectum and anal canal
751.2
Renal agenesis and dysgenesis
753.0
Reduction of limb
755.2–755.4
Anomalies of abdominal wall
756.7
Down’s syndrome
758.0
Table 1 summarizes the procedures carried out on all
hospitalizations for congenital anomalies in the
MED-ECHO database involving newborns and infants
under the age of one.
The data on stillbirths were taken from the stillbirths
database for the calendar years 1989–1996. No melding
is needed for stillbirths because the cases are unique.
Prevalence and Comparisons
Once we had finished melding the Quebec data on
congenital anomalies, we estimated the prevalence of
particular congenital anomalies and compared our results
with the CCASS results for Canada and Quebec. We also
used our Quebec data to examine changes over time and,
for certain congenital anomalies, the disparities among
health and social services regions.
Like the LCDC researchers, we selected 14 major,
relatively common and fairly easily diagnosed congenital
anomalies in order to compare our data with the CCASS
data5 (Table 2).
The Quebec rates were calculated for the period
1989–1991 for each of the anomalies and compared with
the CCASS rates for the same period.5 We also
53
compared the rates for the 1989–1991 period with those
for 1993–1995. Finally, for the 10 most common
anomalies among the 14 selected, we examined regional
disparities during 1989–1995.
The comparison of MED-ECHO data over time shows
little variation (Table 4). Between 1989–1991 and
1993–1995, only the rate of anomalies of the abdominal
wall increased significantly, and only the spina bifida
rate declined.
Results
The analysis of regional data for 1989–1995 (Table 5)
presents some limitations, given the strong variability of
the data as represented by the coefficient of variation. In
six cases, the rates are significantly above the Quebec
average: spina bifida in Saguenay–Lac-Saint-Jean and
the Terres-Cries-de-la-Baie-James region; congenital
hydrocephalus in the Côte-Nord region; congenital
atresia and stenosis of the large intestine, rectum and
anal canal in Bas-Saint-Laurent; anomalies of the
abdominal wall in Estrie; and Down’s syndrome in
Montréal-Centre. In nine cases, the prevalence of a
particular congenital anomaly is significantly lower than
in Quebec as a whole.
During 1989–1991, the CCASS data showed that
Quebec recorded prevalences significantly higher than
the Canadian average for 9 out of 14 anomalies (Table
3). However, the results were obtained from data
covering only 15% of births throughout Quebec and
therefore were not necessarily representative of the
Quebec situation. In the 1995 Status Report on CCASS,
the authors explained this data limitation: “Since
hospitals report to HMRI on an individual basis,
hospitals that choose to participate may be more
specialized and receive more readmissions of infants
with congenital anomalies. This will result in higher
provincial rates being reported.”5
Discussion
The prevalence rates estimated from the MED-ECHO
data, which cover almost all births in Quebec, present
another picture entirely. Quebec does not have a
significantly higher prevalence than the Canadian
average for any anomaly; on the contrary, Quebec’s rates
are significantly lower than Canada’s for five anomalies.
The CCASS data for Quebec are taken from the
records of the HMRI, which covers only a small
proportion of Quebec births and therefore cannot be used
to accurately estimate the prevalence of congenital
anomalies in Quebec.5
TABLE 3
Prevalence rates (per 10,000 total births) of particular congenital anomalies (in
infants <1 year old) by data source, 1989–1991, Quebec and Canada
Data source
ICD-9 code
Congenital anomaly
5
CCASS:
Quebec
MED-ECHO and
stillbirths database:
Quebec
740.0–740.2
Anencephalus and similar anomalies
741.0–741.9
Spina bifida
742.0
Encephalocele
742.3
Congenital hydrocephalus
6.8
14.6
(+)
7.7
745.1
Transposition of great vessels
4.9
15.3
(+)
4.8
746.7
Hypoplastic left heart syndrome
2.6
8.0
(+)
3.4
749.0
Cleft palate
6.9
11.8
(+)
7.3
749.2
Cleft palate with cleft lip
4.9
750.3
Tracheo-esophageal fistula, esophageal
atresia and stenosis
3.4
751.2
Atresia and stenosis of large intestine,
rectum and anal canal
5.4
753.0
Renal agenesis and dysgenesis
5.2
755.2–755.4
Reduction of limb
5.3
756.7
Anomalies of abdominal wall
4.5
758.0
Down’s syndrome
*
1.1
(-)
5
CCASS:
Canada
**
6.8
*
1.0
12.4
(-)
(-)
(-)
(-)
**
*
1.4
2.4
9.9
7.8
1.6
1.5
*
7.8
*
8.5
(+)
8.2
3.8
13.4
(+)
5.8
*
7.5
(+)
5.0
*
5.2
*
7.3
(+)
4.7
24.9
(+)
14.3
4.6
* Coefficient of variation greater than 16.5% and less than or equal to 33.3%. The value should be interpreted with caution.
**
Coefficient of variation greater than 33.3%. The value is shown as an indication only.
(+)/ (-)
Rate significantly higher or lower than the Canadian rate (p # 0.05)
54
Chronic Diseases in Canada
Vol 20, No 2
TABLE 4
Number of cases and prevalence rates (per 10,000 total births) of particular congenital
anomalies (in infants <1 year old), 1989–1991 and 1993–1995, Quebec
ICD-9 code
Congenital anomaly
740.0–740.2
Anencephalus and similar
anomalies
741.0–741.9
Spina bifida
742.0
Encephalocele
742.3
Number of
cases:
1989–1991
Number of
cases:
1993–1995
Rates:
1989–1991
Rates:
1993–1995
40
23
* 1.1
* 0.9
194
147
6.8
5.5
28
32
* 1.0
* 1.2
Congenital hydrocephalus
193
183
6.8
6.9
745.1
Transpositon of great vessels
140
139
4.9
5.2
746.7
Hypoplastic left heart syndrome
75
73
2.6
2.7
749.0
Cleft palate
195
195
6.9
7.3
749.2
Cleft palate with cleft lip
140
148
4.9
5.6
750.3
Tracheo-esophageal fistula,
esophageal atresia and stenosis
97
91
3.4
3.4
751.2
Atresia and stenosis of large
intestine, rectum and anal canal
153
137
5.4
5.1
753.0
Renal agenesis and dysgenesis
148
164
5.2
6.2
755.2–755.4
Reduction of limb
150
124
5.3
4.7
756.7
Anomalies of abdominal wall
127
171
4.5
6.4
758.0
Down’s syndrome
353
332
12.4
12.5
Variation
from
1989–1991
to
1993–1995
–
•
* Coefficient of variation greater than 16.5% and less than or equal to 33.3%. The value should be interpreted with caution.
(–) / (•) Significant reduction or increase between the two periods (p # 0.05)
This study shows that using hospitalization data from
MED-ECHO could be a worthwhile solution for
CCASS. The MED-ECHO database covers virtually all
Quebec births and contains the necessary information for
melding the various hospitalizations for one individual.
The study makes it possible for the first time to
accurately estimate the prevalence of particular
congenital anomalies in Quebec and follow them over
time. It also enables regional comparisons to be made
and Quebec rates to be compared with overall rates in
Canada.
The results show that, for the selected anomalies, the
prevalence rates tend to be lower in Quebec than in
Canada and there is little variation over time. Because of
the strong variability of the data measured by region,
regional comparisons do not yield clear trends.
To obtain even more accurate data on congenital
anomalies, it will be necessary to include the information
on medical termination of pregnancy for congenital
anomalies, and on miscarriages and abortions, as has
been done in international systems.4,11,12 Such additions
1999
are especially worthwhile given the improvement and
greater availability of early detection methods
(ultrasonography, cord puncture, amniocentesis,
trophoblast biopsy, etc.).13,14
Although the deterministic approach is superior to the
probabilistic approach for linking the various events
relating to one individual, using a single identifier to
follow one person among all the data sources is by far
the most accurate method of measuring the actual
number of congenital anomalies.8,10
The role played by administrative practices in relation
to hospitalization should be measured in geographic and
temporal comparisons of the prevalence of congenital
anomalies. It is possible that higher levels of congenital
anomalies simply reflect a greater propensity to indicate
a congenital anomaly code.
Finally, it remains to be seen which of the anomalies
could be selected as sentinel causes.
55
TABLE 5
Prevalence rates (per 10,000 total births) of particular congenital anomalies by region, Quebec, 1989–1995
ICD-9 code
741.0–741.9
742.3
745.1
749.0
749.2
751.2
753.0
755.2–755.4
756.7
758.0
Region Spina bifida
Congenital
hydrocephalus
Transposition of
great
vessels
Cleft palate
Cleft
palate
with cleft
lip
Atresia and
stenosis
of large
intestine,
rectum and
anal canal
Renal
agenesis
and dysgenesis
Reduction
of limb
Anomalies of
abdominal
wall
Down’s
syndrome
01
*
*
*
3.1
*
*
3.8
*
9.4
*
*
1.9
*
02
*
10.6 (+ *
)
7.1
*
5.5
*
11.0
*
7.5
*
5.5
03
*
9.1
*
4.9
9.3
*
5.5
*
5.9
5.7
5.5
*
15.1 (+ *
)
*
5.7
*
7.5
*
*
5.0
*
11.9
9.0
*
7.1
*
5.9
*
10.6
7.7
*
3.9
*
3.9
*
5.3
*
8.6 (-)
*
9.4 (+ *
)
8.2 (-)
04
*
6.6
*
8.1
*
3.3
*
7.9
*
6.3
*
3.8
*
5.1
*
2.8 (-)
05
*
4.9
*
7.8
*
5.3
*
7.4
*
4.9
*
6.6
*
*
2.9
*
4.1
06
4.1 (-)
6.8
5.2 (-)
5.0
4.4
5.0
5.5
4.5
12.0
15.9 (+
)
4.9
07
*
7.4
*
*
1.6
*
4.1
*
4.1 (-)
*
6.2
*
*
2.5
*
*
2.9
*
3.7
*
*
2.1
*
9.4
08
*
9.6
*
*
5.1
*
*
3.8
*
8.9
*
6.4
*
5.7
*
5.7
*
*
3.8
*
6.4
*
7.7 (-)
09
*
10.1
*
14.2 (+ *
)
*
1.0
*
10.1
*
*
8.1
*
*
3.0
*
*
7.1
*
*
6.1
*
*
8.1
*
13.2
10
*
*
4.4
*
*
4.4
*
*
0.0
*
*
4.4
*
*
13.3
*
*
4.4
*
*
0.0
*
*
4.4
*
*
4.4
*
*
4.4
11
*
12.5
*
*
7.5
*
*
7.5
*
*
7.5
*
*
1.2
*
*
3.7
*
*
3.7
*
*
5.0
*
*
3.7
*
22.4
12
*
8.1
*
6.9
*
5.7
*
5.4
*
6.0
*
6.6
*
7.5
*
4.8
*
6.6
15.6
*
3.2 (-)
*
3.5
*
7.1
*
5.1
*
4.2
*
5.8
*
5.1
*
*
1.9
13.1
13
*
5.4
14
*
2.5 (-)
*
5.6
*
5.6
*
6.4
*
4.2
*
5.8
*
6.9
*
7.5
*
6.7
11.7
15
*
5.7
*
5.5
*
3.2
*
4.7
*
4.5
*
5.5
*
5.7
*
3.5
*
4.0
10.5
17
*
*
0.0
*
*
16.7
*
*
0.0
*
*
5.6
*
*
44.5
*
*
27.8
*
*
0.0
*
*
5.6
*
*
22.2
*
*
22.2
18
*
44.3 (+ *
)
*
9.9
*
*
4.9
*
*
19.7
*
*
4.9
*
*
0.0
*
*
4.9
*
*
4.9
*
*
9.9
*
*
19.7
16
5.1
6.3
5.0
7.1
5.5
4.1
4.
References
1.
Ministère de la Santé et des Services sociaux. Politique de
périnatalité. Quebec: Government of Quebec, 1993:21–2.
2.
Ministère de la Santé et des Services sociaux. La politique
de la santé et du bien-être. Quebec: Government of
Quebec, 1992:66–71.
3.
Soltani MS, Guediche MN, Bchir A, Ghanem H, Pousse H,
Braham A. Facteurs associés aux faibles poids de
56
Chronic Diseases in Canada
5.0
6.6
11.0
naissance dans le Sahel tunisien. Archives françaises
pédiatriques 1991;48:405–8.
Acknowledgements
This study received financial support from the Bureau of
Reproductive and Child Health of the Laboratory Centre for
Disease Control. The authors would like to thank Jocelyn
Rouleau of that Bureau.
5.6
5.
6.
De Wals P, Royer-Trochet C. Évaluation des bases de
données sanitaires pour la recherche et la surveillance
épidémiologique des malformations congénitales. Régie
régionale de la santé et des services sociaux de la
Montérégie, 1997.
Rouleau J, Arbuckle TE, Johnson KC, Sherman
JG. Description and limitations of the Canadian Congenital
Anomalies Surveillance System (CCASS) [status report].
Chronic Dis Can 1995;16(1):37–42.
Johnson KC, Rouleau J. Temporal trends in Canadian birth
defects birth prevalences, 1979–1993. Can J Public Health
1997;88(3):169–76.
Vol 20, No 2
7.
8.
Ministère de la Santé et des Services sociaux. Les
banques de données du MSSS. Numéro 1. Données sur la
clientèle hospitalière (MED-ÉCHO). Quebec: MSSS,
1986.
Turner D, Roos LL, Traverse D, Stranc L, Harrison M,
Fields ALA, Bryant H. Forging partnerships through data:
a general strategy for record linkage. In: Information
technology in community health. Victoria: School of
Health Information Science, University of Victoria;
1998:1-2–1-6.
9. Canadian Congenital Anomalies Surveillance System.
Melding process description [working document].
Ottawa: Laboratory Centre for Disease Control, 1997.
10. Roos LL, Wajda A. Record linkage strategies. Part I:
estimating information and evaluating approaches.
Methods of Information in Medicine 1991;30:117–23.
1999
11. EUROCAT. 15 years of surveillance of congenital
anomalies in Europe 1980–1994. Brussels: Scientific
Institute of Public Health - Louis Pasteur, 1997.
12. International Clearinghouse for Birth Defects Monitoring
Systems. Congenital malformations worldwide.
Amsterdam, 1991.
13. Gérard CH, Gillerot Y, Koulischer L, Hustin J.
Amniocentèse et biopsie trophoblastique. Journal
gynécologique-obstétrique-biologique et reproductif de
Paris 1991;20:617–22.
14. Gougard J, Ayme S, Stoll CL. L’évaluation des
technologies diagnostiques des malformations
congénitales. Journal gynécologique-obstétriquebiologique et reproductif de Paris 1992;21(3):278–80. O
57
Monograph Series on Aging-related Diseases:
XII. Parkinson’s Disease—Recent Developments
and New Directions
Natalie Kontakos and Julie Stokes
Abstract
Parkinson’s disease, a chronic progressive disorder of the central nervous system
characterized by tremor, rigidity and bradykinesia, usually affects those over the age of 50.
Recent developments in research on Parkinson’s disease include investigation of the possible
role of diet and a growing interest in genetics and inherited factors. The identification of
biological markers and other environmental risk factors will play a significant role in future
research of the disease as they will be important in the development of prevention strategies.
Key words: Canada; diagnosis; morbidity; mortality; risk factors; treatment
Introduction
Parkinson’s disease (PD) is a neurodegenerative
disorder that primarily affects voluntary, co-ordinated
movement. It is a disease of late middle age, usually
affecting those over the age of 50. Although the
discovery of PD is often attributed to James Parkinson
and his 1817 monograph entitled The Shaking Palsy,1
descriptions of parkinsonian syndromes date back to the
ancient Ayurvedic literature of India, from 4500 to1000
BC.2 The first breakthrough in PD research was in the
1960s, when the dopamine hypothesis and levodopa
therapy were introduced.1 There has since been much
progress in disease definition and diagnosis,
surveillance, knowledge of etiology and disease
progression, and treatment. Although the cause of PD is
not yet known and a cure has not been found, the past
few years of research have lead to a greater
understanding of the disease. As well as providing an
overview of PD, this report focuses on the recent
advances and the future directions of PD research.
Background and Natural History
PD is a chronic and progressive disorder of the central
nervous system. It is the most common form of the
parkinsonian syndromes, a group of motor system
disorders that share the primary symptoms of tremor,
rigidity and bradykinesia. Most studies indicate that
there must be two of these three features for a diagnosis
of parkinsonism.3–5
Parkinsonian syndromes occur when the neurons that
lie in the brain stem’s substantia nigra (“black
substance”) are destroyed.4 The neurotransmitter
dopamine is normally produced in the neurons of the
substantia nigra. These neurons connect with other
neurons in the corpus striatum, which in turn send
messages to the motor-controlling areas of the cortex.
Dopamine is depleted as the neurons of the substantia
nigra diminish in number; therefore, the number of
signals to the corpus striatum and from there to the
cortex are decreased. The normal functioning of the
motor system is thus disrupted (Figure 1).
Depletion of dopamine in the brain can come about in
a number of ways. Parkinsonian syndromes may be
induced by drugs, viral infections, hereditary diseases or
metabolic causes. Parkinsonism and syndromes such as
progressive supranuclear palsy and multiple system
atrophy may present as relatively pure parkinsonism in
the early stages of disease, with nonparkinsonian signs
becoming more prominent with time. Other degenerative
diseases of the central nervous system may either occur
concurrently with PD or may exhibit some parkinsonian
Author References
Natalie Kontakos and Julie Stokes, Aging-related Diseases Division, Bureau of Cardio-Respiratory Diseases and Diabetes, Laboratory Centre for
Disease Control, Health Protection Branch, Health Canada (funding provided by Division of Aging and Seniors, Population Health Directorate,
Health Promotion and Programs Branch, Health Canada)
Correspondence: Julie Stokes, LCDC Building, Health Canada, Tunney’s Pasture, Address Locator: 0602E2, Ottawa, Ontario K1A 0L2
58
Chronic Diseases in Canada
Vol 20, No 2
symptoms. Stroke, tumours, trauma and
other nondegenerative conditions may
influence the level of dopamine in the
substantia nigra and/or corpus striatum
and thus include parkinsonian features.
FIGURE 1
Difference between a normal brain and a parkinsonian brain
PD is different from most other
parkinsonisms in that the cause of the
destruction of the substantia nigra
leading to dopamine reduction is not
known.4 Many ideas, such as the
“oxidative stress model”,6,7 have been
brought forward to explain how PD
begins, but none has been fully accepted.
The pathological classification of PD
includes the degeneration of specific
groups of nerve cells, including the substantia nigra.4
Poor circulation or arteriosclerosis cannot explain the
location of the affected cells. PD can also be diagnosed
after death on the basis of the Lewy body, a round
inclusion found within degenerating neurons. Lewy
bodies are highly characteristic of the disease.
Burden of Disease
Morbidity
According to a recent World Health Report, PD
affects 3,765,000 individuals worldwide, and the
condition is diagnosed in 305,000 people per year.8 In
1996, there were 2,635,000 people with PD who were
chronically disabled and 58,000 deaths. Although PD
affects individuals worldwide in all ethnic groups and
from all socio-economic backgrounds, statistics
reflecting the disease’s morbidity and mortality vary
widely from place to place. In fact, a recent review of the
worldwide occurrence of PD9 revealed that there was a
13-fold difference between the highest (Uruguay) and
the lowest (China) prevalence estimates in door-to-door
studies, and a 3-fold difference between the highest
(Iceland) and the lowest (Libya) locations in studies
relying on data from sources such as hospitals,
physicians and health insurance records. Incidence
estimates exhibited a 10-fold difference between the
highest (United States) and lowest (China) areas.
These ranges in prevalence and incidence may
suggest environmental or genetic clues to the disease’s
etiology. They may also be due to other factors, such as
differences in diagnostic procedures or population
groups. Since this review standardized the rates for all
studies to a single population, the variations cannot be
attributed to populations of different age structures. Case
ascertainment may be used to explain the difference in
estimates between door-to-door studies and other studies
that do not actively seek out individuals with PD.
The literature in the past has been quite consistent in
reporting higher PD rates in primarily Caucasian
populations as compared with Asian or black
1999
populations. More recent studies indicate that the
variation in the prevalence of PD among different ethnic
groups is not as large as it was once thought to be, but
prevalence still varies from study to study.10–17 A greater
consistency in study methodologies probably explains
this shift.
Recent incidence studies conducted in the United
States and Europe all reveal PD incidence rates between
8 and 13 per 100,000.15,18–20 All studies that included
information specific for each sex reported higher
incidence rates among males than females. The high
male incidence rate in a study completed in Manhattan,
United States,15 is largely attributed to the incidence rate
among black males; in comparison to the rate among
white males it was consistently higher in every age
group, with a 4-fold difference in those over the age of
80. The duration of disease is probably shorter in black
males, since the same study observed lower prevalence
rates among black as compared with white males.
A Swedish study that used records from a health
maintenance organization (HMO)20 observed a higher
crude incidence rate among whites than among blacks,
Hispanics or Asians; Asians had the lowest rate. A
cohort of male Hawaiian residents, of primarily Japanese
or Okinawan ancestry,18 had an incidence rate in
between that of the Manhattan study and the HMO
study. The authors concluded that environmental factors
were more important than genetic factors in this group of
men, since Asian incidence rates reported previously
were lower.
All studies showed incidence rates that increased
linearly with age up until the age of 75. At this point, the
incidence rate in most groups either plateaued or
continued to increase linearly. The incidence rate in the
Hawaiian cohort, however, decreased, and the rate
among males in Manhattan increased even more steeply.
Morbidity in Canada
There are few data describing the prevalence and
incidence of PD in Canada. Since most patients do not
require hospital care on an in-patient basis, hospital
59
estimate the prevalence of PD in Canada. A 1988 British
Columbia rural community study21 revealed a crude
prevalence rate of 69 per 100,000, which is considered to
be low compared with that of other communities. No
age-specific rates were reported, however, so
comparisons could not be made. Another study involved
a cohort of individuals registered with the Alberta Health
Care Insurance Plan who were followed for a five-year
period.22 PD patients were identified through physician
billing information in which a diagnostic code for PD
was included. The crude prevalence rate was found to be
244.4 per 100,000. The rate was higher among males
than females, and 81% of all cases were 60 years of age
or older.
separation rates underestimate the prevalence of PD.
These rates are also problematic in that they are not
based on the number of individuals but, rather, on the
number of discharges. An individual can therefore be
counted more than once. Despite the drawbacks, hospital
separation rates can be useful in detecting general
differences and trends. The overall hospital separation
rate for PD (coded as paralysis agitans ICD-9 332.0) for
the 1991–1995 period was 15.4 per 100,000 among men
and 8.9 per 100,000 among women (Table 1). There was
considerable variation from province to province,
ranging from 8.0 per 100,000 in Newfoundland to 19.3
per 100,000 in Saskatchewan. This variation may be
partly explained by different disease coding or hospital
admission practices.
A study conducted in Saskatchewan found that 3% of
individuals over the age of 65 had PD.23 This estimate is
rather unstable, since it was based on only two positive
cases among 70 subjects. A similar study found a 6%
rate of PD in a chronic care facility.24 Studies in other
countries have revealed that the higher prevalence of PD
among those living in chronic care facilities is largely
due to a higher prevalence in the “young-old” age
groups.25,26
During 1991S1995, hospital separation rates increased
with age and peaked among individuals aged 80–84.
Males had higher hospitalization rates than females in
every age category (Table 2), and the overall hospital
separation rates were higher among males in every
province.
From 1976S1980 to 1991S1995, there was a 25%
reduction in hospital separations for PD over all age
groups. Among females the rates have been decreasing
since 1976S1980, but the rates among males peaked
during 1986S1990. The decline in hospital separation
rates can be attributed to the fall in the younger age
groups, especially in females. There has actually been an
increase in hospital separation rates in the older age
groups.
Mortality
International mortality rates increase with age and are
consistently higher among males. Recently published
mortality rates show that rates are similar in European
countries27–29 and lower in Japan.30 There has been a
steady increase in mortality rates among older
populations (>75 years) and declining rates among
younger populations (<65 years).31
Excluding routine data such as hospital separation
rates, there have been very few documented attempts to
TABLE 1
Average annual hospital separation ratesa (per 100,000) for Parkinson’s disease by sex, province and
period, Canada, 1976–1995
1976–1980
Province
1981–1985
1986–1990
1991–1995
Males
Females
Males
Females
Males
Females
Nfld
17.2
10.5
14.4
10.7
14.2
10.3
9.3
7.2
PEI
13.2
18.3
15.6
21.2
11.2
15.0
12.2
7.3
NS
14.2
9.5
16.3
11.0
17.1
10.7
12.8
6.7
NB
20.1
13.2
18.4
13.3
17.1
10.5
15.2
9.1
Que
7.9
6.7
9.4
6.6
12.7
7.5
13.6
8.2
Ont
19.2
13.7
19.2
11.6
19.4
10.4
14.1
7.8
Man
20.0
14.0
19.1
14.0
21.9
12.6
19.9
11.4
Sask
31.1
24.9
27.0
23.1
32.4
20.6
23.5
16.4
Alta
27.4
22.1
30.4
21.5
22.7
14.2
12.0
8.0
BC
23.8
16.7
26.3
15.8
26.5
16.4
21.2
11.9
CANADA
18.3
13.2
19.0
12.2
19.6
11.3
15.4
8.9
a
60
Males
Females
Standardized to the 1991 census population
Source: Laboratory Centre for Disease Control, based on data from Statistics Canada
Chronic Diseases in Canada
Vol 20, No 2
TABLE 2
Average annual hospital separation ratesa (per 100,000) for Parkinson’s disease
by sex, age and period, Canada, 1976–1995
1976–1980
Age
(years)
Males
1981–1985
Females
1986–1990
1991–1995
Males
Females
Males
Females
Males
Females
ALL AGES
18.3
13.2
19.0
12.2
19.6
11.3
15.4
8.9
45–64
14.5
12.2
13.4
9.4
10.7
7.4
7.9
6.2
32.2
65–69
66.3
61.8
62.6
51.6
62.7
39.3
43.2
70–74
133.9
104.1
125.0
92.3
126.0
85.2
95.1
58.2
75–79
176.7
124.6
208.7
131.6
206.0
129.8
169.0
100.9
80–84
210.3
116.8
250.8
131.4
293.9
137.5
232.4
117.2
85+
188.1
78.6
200.2
87.9
258.8
96.8
224.4
84.5
Mortality in Canada
Although most PD patients do not die as a result of
the disease, mortality data can be examined to identify
differences in the disease’s distribution according to
geographic area, sex, age and time. Mortality rates may
also draw attention to differences in treatment and
management. The overall mortality rate for PD during
the 1992–1996 period was 3.4 per 100,000 (Table 3).
Two provinces had mortality rates that were significantly
different from the national rate: Ontario showed a
significantly higher rate (3.7 per 100,000) and Alberta a
significantly lower rate (2.6 per 100,000) for all ages. As
with other rates, these differences may be real or they
may be due to other factors, such as provincial
differences in coding death certificates.
During 1992S1996, mortality rates increased with age
and did not reach a peak like the hospital separation rates
(Table 4). Males had higher mortality rates than females
in all age groups, and, as with hospital separation rates,
overall mortality rates were higher among males in all
provinces.
Over time, standardized mortality rates have increased
among both males and females. The increase among
males from 1977S1981 to 1992S1996 was greater (93%)
than the increase among females (79%). As with hospital
separation rates, the increase in PD mortality is largely
attributed to a greater increase of PD in older age groups
than in younger age groups. The mortality rates in
younger age groups, however, have not decreased to the
extent that hospital separation rates have.
TABLE 3
Average annual mortality ratesa (per 100,000) for Parkinson’s disease by
sex, province and period, Canada, 1977–1996
1977–1981
Province
Males
1982–1986
Females
Males
1987–1991
Females
Males
1992–1996
Females
Males
Females
Nfld
1.8
0.9
2.1
2.3
5.6
1.9
5.4
2.7
PEI
1.5
0.8
3.0
1.4
4.7
2.6
2.8
2.0
NS
2.3
0.7
2.5
1.4
4.1
1.8
4.4
2.1
NB
2.2
1.4
2.5
1.8
3.5
1.7
4.4
2.1
Que
2.1
1.1
2.6
1.5
4.0
2.1
4.9
2.5
Ont
2.6
1.4
3.8
1.7
4.7
2.2
5.7
2.6
Man
2.5
1.2
3.4
1.4
3.8
2.2
5.1
2.0
Sask
2.6
1.1
2.6
1.5
2.8
1.9
4.9
2.2
Alta
2.2
1.7
3.0
1.7
3.2
1.7
3.8
1.8
BC
4.3
2.0
3.0
1.7
4.3
1.9
4.7
2.5
CANADA
2.6
1.4
3.1
1.6
4.2
2.0
5.1
2.4
a
Standardized to the 1991 census population
Source: Laboratory Centre for Disease Control, based on data from Statistics Canada
1999
61
TABLE 4
Average annual mortality ratesa (per 100,000) for Parkinson’s disease
by sex, age and period, Canada, 1977–1996
1977–1981
Age
(years)
Males
1982–1986
Females
Males
1987–1991
Females
1992–1996
Males
Females
Males
Females
ALL AGES
2.6
1.4
3.1
1.6
4.2
2.0
5.1
2.4
45–64
0.7
0.4
0.6
0.3
0.6
0.3
0.6
0.4
65–69
5.8
3.1
5.8
3.2
6.0
2.6
6.2
2.8
70–74
16.1
6.7
16.0
7.6
17.4
7.9
19.1
9.8
75–79
28.7
14.4
35.2
20.4
42.8
21.0
56.2
23.8
80–84
43.5
24.1
60.0
31.8
81.1
40.2
92.4
47.1
85+
53.6
30.2
72.0
34.6
123.4
62.2
158.2
77.4
a
Standardized to the 1991 census population
Source: Laboratory Centre for Disease Control, based on data from Statistics Canada
Risk Factors
Environmental Factors
The search for an environmental agent causing PD
has been quite intensive. It heightened in the mid to
late 1980s when MPTP (1-methyl-4-phenyl-1,2,3,6tetrahydropyridine), a rare contaminant of heroin, was
found to elicit clinical and pathological features virtually
identical to those of PD.32,33 It was thought that an
environmental toxin with chemical and physical
properties similar to MPTP could lead to PD. Although
no such toxin has been found to be causally associated
with PD, studies have offered substantial evidence to
eliminate certain hypotheses and to explore other
hypotheses further.
Rural living
Although studies in the past had pointed to an
association between rural living and PD, the most recent
studies are quite inconsistent in their results. Not only
are the results inconsistent, but so are the periods of
exposure under investigation.
Two studies found elevated and significant odds ratios
(ORs) for rural living near the time of diagnosis,34,35
whereas another study found PD mortality to be
positively correlated with population density.36 A
Chinese study that found an OR of less than 1 for living
in small cities did not specify the exposure period of
interest.37 One study from the US found an elevated and
significant OR for a history of rural residence only for
blacks and not for other ethnic groups.38 The average
population density of places of residence from birth to
the time of diagnosis did not differ between cases and
controls in one study,39 and another showed no
association between the place of residence for the first 15
years of life and the risk of PD.40 Since PD is believed to
have a long latent and asymptomatic period,41 the
62
Chronic Diseases in Canada
probable relevant exposure period would be earlier rather
than later in life.
Because there are many specific exposures associated
with rural areas, recent studies have attempted to
measure these exposures in order to explain the
association between PD and areas of low population
density.
Three recent studies36,42,43 have found an association
between PD and agricultural work. One of them43 looked
solely at death certificates, and another36 found a positive
correlation between the number of PD deaths and
farming density. In the third study,42 the OR for grain
and crop farming was significant in univariate analysis
but not in multivariate analysis. A German study
reported an elevated and significant relative risk
associated with mushroom harvesting during childhood
and adolescence;44 however, no association was found
with previous farm activity or employment in
agricultural work, living on a farm or having a farm near
to the home, having contact with farm animals or
involvement in slaughter. Other recent studies34,38 have
found no association between PD risk and farming.
Pesticide exposure
The similarity between the structures of the MPTP
metabolite MPP+ and the herbicide paraquat encouraged
the investigation of a possible relation between pesticide
exposure and PD (Table 5).34,35,39,45–48 A study in Taiwan
indicates that the OR for PD was 2.0 among those
subjects who had used both paraquat and other
herbicides/pesticides when compared with those exposed
to pesticides and herbicides other than paraquat.48 Recent
studies have consistently shown an increased risk of PD
with pesticide exposure, and in some this has achieved
statistical significance.
Vol 20, No 2
Well water
Well water has also been implicated in PD.
Of all the recent studies, only those conducted
in Italy49 and Spain50 have found a positive
association between drinking well water and
the risk of PD. A study that examined the
relation between PD mortality and the
proportion of well water users in Michigan
showed a negative correlation between the
exposure and the disease,36 and a Chinese
study also reported a decreased risk of PD
associated with drinking well water.37 Four
studies found no association,38–40,48 and no
contaminant thought to contribute to the
cause of PD has been found in well water.
TABLE 5
Recent studies investigating the association between
pesticide use and risk of Parkinson’s disease (PD)
Study authors
Location
Relation between pesticide use and
PD risk
Butterfield et al.34
United States
Insecticide exposure: odds ratio (OR) = 5.75
(significant)
Past residency in a fumigated household:
OR = 5.25 (significant)
Herbicide exposure: OR = 3.22 (significant)
Hertzman et al.47
Canada
Occupational pesticide use in males: OR =
2.03 (95% CI = 1.00–4.12) versus cardiac
disease controls and OR = 2.32 (95%
confidence interval [CI] = 1.10–4.88) versus
electoral list controls
Not significant in women
United States
Pesticide use: OR = 3.42 (95% CI =
1.27–7.32)
Taiwan
Occupational or residential exposure to
herbicides/pesticides: OR = 2.89 (95% CI =
2.28–3.66)
Paraquat exposure: OR = 3.22 (95% CI =
2.41–4.31)
Spain
More cases than controls were exposed to
pesticides but relative risk estimate was not
significant
Germany
Herbicide use: significant for 1-dose category
versus regional controls (41–80 dose-years:
OR = 3.0 [95% CI = 1.5–6.0]) but not versus
neighbourhood controls
Insecticide use: significant OR in the 2 lower
dose categories versus regional controls; OR
not significant versus neighbourhood controls
Organochlorines: significant OR versus
regional controls
Alkylated phosphates and carbamates:
significant OR versus regional controls
Metal exposure
Some metals, such as manganese and
Hubble et al.46
mercury, have been shown to induce
parkinsonian signs and symptoms in
Liou et al.48
individuals who were heavily and acutely
exposed.51 Recent epidemiologic studies have
looked at mainly occupational metal exposure
as a risk factor for PD. No significant ORs
Morano et al.35
were found in studies conducted in British
Columbia47 or Alberta.45 A German study
found only one significant OR for
Seidler et al.39
occupational exposure to lead.39 This,
however, was barely significant and only
when compared with one of two control
groups. Another OR that was just significant
and involved only one control group was
found for exposure to mercury through
amalgam fillings. An ecological study
reported that counties in Michigan with iron
and copper industries had higher PD mortality
rates.36 In another study, elevated ORs for 20
years of occupational exposure to copper,
manganese and various combinations of
metals suggested that metal exposure may
play a role in PD etiology.52 However, the
finding that serum and urine levels of
manganese, chromium and cobalt did not differ between
PD patients and controls led the authors of another study
to suggest that exposure to these metals is unrelated to
PD.53
Non-metallic toxins
The relation between PD and numerous other toxins
has been investigated. Although one study54 reported
positive associations with plastic resins, epoxy resins,
glues, paints and petroleum, it also found a multitude of
other exposures not to be significant, leading to a
problem with multiple comparisons. Higher mortality
rates from PD were found in counties with paper and
chemical industries.36 Other studies have not found a
relation with industrial toxins,35 carbon monoxide,39,45
cyanide,45 exhaust fumes39 or glues, paints and
1999
lacquers.39 More PD patients than control patients have
reported that they had wood panelling in their homes;55
this may implicate wood preservatives in the etiology of
PD.
Head injury
Head injury has been implicated in the etiology of
PD, possibly through the microglial cells, which are
involved in the inflammatory process.6 Two of
four35,39,45,56 recent studies found significant ORs with
head trauma. The OR in one of these was barely
significant and existed only in a comparison with one
control group;39 the other reported an OR that remained
significant after multivariate analysis.45 Studies
examining head trauma may involve recall bias, and this
issue should be addressed in future studies.
63
Smoking
Epidemiologic studies have consistently shown
smoking to be protective for PD. The majority of recent
studies seem to support this relation, as they have
reported ORs of less than 1 (Table 6).34,39,40,45,56–59 In
addition to this evidence, a prospective study involving
the Honolulu Heart Study revealed a significant relative
risk of 0.39.60 Some experimental evidence supports the
idea that nicotine may be protective for PD. One study
showed that chronic nicotine intake in rats decelerated
the age-associated decrease in dopamine receptors and in
dopamine re-uptake.61
Other case-control studies, however, do not support
the claim that smoking is protective for PD. Longitudinal
Gompertzian analysis, which considers the three
dimensions of genetics, environment and selective early
mortality, demonstrates that a neuroprotective influence
does not explain the negative association between PD
and smoking.62 The negative association has been
explained by the fact that smokers die sooner than
non-smokers.
TABLE 6
Recent case-control studies examining the
relation between smoking and Parkinson’s
disease (PD)
Study authors
Location
OR
Comments
United States
0.50*
0.43*
0.37*
At 5 years before diagnosis
At 10 years before diagnosis
At 15 years before diagnosis
Hellenbrand et al.59
Germany
0.5*
0.2*
History of smoking
Current smoker
Martyn and
Osmond56
England
0.50*
History of smoking
Mayeux et al.57
United States
1.1
0.20*
History of smoking
At the time of interview
Seidler et al.39
Germany
—
PD patients reported fewer
pack-years
Semchuk et al.45
Canada
0.48*
0.58
Univariate analysis
Multivariate analysis
Tzourio et al.58
Europe
(France, Italy,
Spain,
Netherlands)
1.1
0.4*
History of smoking
<75 years of age and
history of smoking
Vieregge et al.40
Germany
0.37*
0.42
0.24
History of smoking
Smoking for a duration of 2
years
Smoking more than 10
cigarettes per day
Butterfield et
al.34
* p < 0.05
64
Chronic Diseases in Canada
Tzourio et al.58 found no overall protective effect of
smoking in relation to PD but, when adjustments were
made for age, tobacco was found to be protective in the
younger age group while representing an increased risk
in the older age group.
Although smoking has been found to be protective for
PD, it is an important risk factor for many other major
diseases, and the adverse effects of smoking far
outweigh any possible benefits.
Diet
Diet has only recently been implicated in the etiology
of PD. According to the oxidative stress model, an
increase in antioxidants would prevent damage and death
to dopaminergic cells by scavenging more free radicals.
Therefore, antioxidants present in foods and available in
supplements would be protective for PD. Epidemiologic
studies that have examined the association between
antioxidants and PD have been inconsistent (Table
7);63–68 no study replicated any finding of a negative or
positive association. The two prospective studies63,65
used dietary history information collected before PD was
diagnosed; in one,63 the period from dietary to disease
assessment was only a few years.63 Since the disease
process is thought to start many years before the
individual is symptomatic, the dietary information
obtained in the latter study may be irrelevant in terms of
PD etiology. The other three studies64,66,67 did not fare
any better, in that they were case-control studies whose
data focused on the individual’s dietary pattern over the
previous year.
Three other case-control studies34,44,68 examined the
relation between PD and foods rich in vitamin E.
Although there was no difference in intake of foods rich
in vitamin E in two of the studies,44,68 the third reported
an elevated and significant OR for nuts and seeds, which
are rich in vitamin E.34 One of these studies68 found a
higher intake of vitamin C in PD patients.
Studies relating antioxidant serum levels with PD
status have also been performed. Three studies69–71 found
no difference in vitamin E serum levels between PD
patients and healthy controls. One of these studies70 also
found no difference in vitamin A levels but did find
higher vitamin C levels in PD patients; the vitamin C
levels in controls, however, were low compared with
established data in young healthy individuals. It is
important to note that these studies included levels
measured after the time of diagnosis and may not reflect
levels before disease onset.
The relation between other dietary variables and PD
etiology has also been recently examined. An ecological
study reported significant and positive correlations
between age-adjusted mortality rates in 17 different
countries and the per capita consumption of total dietary
protein and meat.72 In addition to the limitations inherent
Vol 20, No 2
TABLE 7
Studies investigating the association between antioxidant intake and risk of Parkinson’s disease (PD)
Study authors
Cerhan et
al.63
Location
Food source
Odds ratio (and 95% CI)
Conclusions/Comments
United States
Vitamin C
(lowest vs highest tertile)
Manganese
Vitamin A
Retinol
Beta carotene, vitamin E, zinc and
selenium
0.5 (0.2–1.0)
Certain antioxidants may be protective
agents for PD while others may be
risk factors
0.4 (0.2–0.9)
2.1 (1.0–4.1)
1.9 (0.9–3.7)
No association
de Rijk et al.67
Netherlands
Vitamin E (/10 mg)
Beta carotene (/1 mg)
Vitamin C (/100 mg)
Flavonoids (/10 mg)
0.5 (0.2–0.9)
0.6 (0.3–1.3)
0.9 (0.4–1.9)
0.9 (0.7–1.2)
Vitamin E may be protective for PD
Gorell et al.66
United States
Vitamins A, B, C, E and beta carotene
No association
No association between intake of
these vitamins and PD
Logroscino et al.64
United States
From supplements:
Carotenoids
Vitamins A, C, E and retinol
From food:
Vitamins
Marginal linear trend (p = 0.095)
Not associated with PD
No difference in antioxidant intake
between PD patients and controls
No association
Morens et al.65
United States
Vitamin E
(continuous variable)
0.88 (0.63–1.23)
Inconclusive results— the possibility
that vitamin E may be protective for PD
is not ruled out
Scheider et al.68
United States
Vitamin E
Vitamin C
Total carotene
1.15 (0.47–2.80)
2.13 (0.89–5.11)
2.27 (0.83–6.17)
No protective effect for vitamin E
Greater PD risk with higher intakes of
vitamin C and carotenoids
in ecologic studies, the study not only used mortality
rates, which in comparison to other statistics do not
accurately indicate the prevalence of PD, but also used
figures from 1952 to 1958, which do not reflect the
present rates. Two case-control studies64,66 found an
elevated and significant OR with fat intake. One of
these66 also reported elevated and significant ORs for
cholesterol, iron and lutein. Since lipids are one of the
major sources of free radicals, the increase in fat and
cholesterol intake is consistent with the oxidative stress
model. The positive association with lutein may be a
result of PD, as many patients increase their
consumption of lutein to manage the disease’s
symptoms.
A German study reported that PD may be related to a
variety of foods.73 PD patients consumed more
chocolate, desserts, organ and raw meats, and less beer
and coffee. The relation between the disease and these
food items might be related to the effects of biogenic
amines (chocolate), insulin levels (foods rich in refined
carbohydrates), infectious agents (organ and raw meats),
ethanol (beer) and caffeine (coffee) on the dopaminergic
system. Another study has also found a negative
association between alcohol consumption and PD.74
1999
Infections
The idea that PD may be infectious in origin is largely
due to the onset of parkinsonian symptoms in individuals
infected with the virus associated with lethargic
encephalitis in the 1920s.75 Numerous studies have failed
to find an association between PD and a variety of
common viruses and bacteria. Chicken pox, measles,
rubella, mumps, the Spanish flu45 and the Nocardia
species76 were all found to be unrelated to PD in recent
studies. A study conducted in the United Kingdom
reported that PD patients were more likely to recall
suffering from croup or diphtheria in childhood.56 It is
important to note however, that these results are not
based on antibody serum levels and that the neurotoxin
produced by the organism of diphtheria cannot cross the
bloodSbrain barrier.
An etiologic hypothesis involving whooping cough
was brought forward when a positive relation between
PD and whooping cough outbreaks in one-year birth
cohorts was found in Iceland.77 PD patients also had
higher antibody responses to coronaviruses than did
healthy controls, suggesting an association between these
RNA-containing viruses and PD.78 The observation that
there is a higher prevalence of gastrointestinal ulcers in
65
PD patients has led to the hypothesis that Helicobacter
pylori, the Gram-negative bacterium responsible for the
majority of cases of ulcers, may have a role in PD
etiology.79
Genetics and Inherited Factors
For many years, epidemiologists focused most of their
attention on environmental risk factors. Hereditary
influences seemed less likely because twin studies in the
past had shown similar concordance rates among
monozygotic and dizygotic twins.55,80,81 More recently,
however, there has been a growing interest in genetic
factors, largely due to the realization that family history
is an important risk factor in the etiology of PD.
Family history
Many epidemiologic studies have investigated the
association between the risk of PD and a family history
of the disease (Table 8).34,35,37,39,45,49,82–85 Most, if not all,
studies have consistently reported a significantly
elevated OR for a family history. Although it is possible
that recall and selection biases may explain some of the
observed association, the consistency, strength and
universality of the results support a role for early life
environmental exposures or some underlying genetic
predisposition to the disease. Furthermore, a study by
Uitti et al.86 identified previously undiagnosed cases of
PD among families who had reported no family history
of the disease, suggesting that patients’ reports of the
absence of familial parkinsonism may be inaccurate. The
results of this study also indicate that the weighted
prevalence rate of familial parkinsonism is more than
five times greater than the reported prevalence rates of
PD in the general population.
A number of families with multiple cases of PD have
been reported in the literature. Some of the most
impressive kindreds include a family with 18 affected
individuals within six generations.87 Not only did
autopsy findings include features consistent with PD, but
clinical symptoms such as age of onset and
responsiveness to levodopa were also in agreement with
typical cases. Other families include the Contursi
kindred with 60 affected individuals in five
generations.88 A Greek-American kindred whose 16
affected members in three generations showed
asymmetric rigidity, resting tremor, bradykinesia and
postural instability were also responsive to levodopa.89
The data from all of these families are consistent with an
autosomal dominant mode of inheritance with reduced
penetrance. In addition, a comparison between familial
and sporadic cases of PD revealed that the clinical
parameters and the course of disease were similar.90
Genetic markers
Many researchers are attributing gene identification in
one family as the biggest breakthrough in PD research
since the observations of dopamine deficiency91 and
subsequent successful symptom control with levodopa.92
66
Chronic Diseases in Canada
TABLE 8
Controlled studies investigating the association
between family history of Parkinson’s disease
(PD) and risk of PD
Study authors
Location
OR
Comments
Bonifati et al.83
Italy
4.95*
Positive family history
Butterfield et al.34
United States
2.97*
Positive family history
De Michele et al.49
Italy
14.6*
Positive family history
Marder et al.85
United States
2.3*
First-degree relatives
Morano et al.35
Spain
3.92*
Positive family history
Payami et al.82
United States
3.5*
First-degree relatives
Seidler et al.39
Germany
12.6*
First- or second-degree
relatives vs
neighbourhood controls
First- or second-degree
relatives vs regional
controls
5.0*
Semchuk et al.45
Canada
2.36*
3.73*
5.76*
First-degree relatives
First- or second-degree
relatives
First-, second- or
third-degree relatives
Multivariate model
Vieregge84
Germany
7.05*
Positive family history
Wang et al.37
China
4.33*
Positive family history
* p < 0.05
An article first reported that genetic markers on
chromosome 4q21-q23 were found to be linked to
individuals with PD in a large Italian kindred.93 Then,
less than one year later, a second article described the
exact gene and mutation thought to be responsible for
PD in this family and other, Greek families.94 A base
pair substitution in the a-synuclein gene was found in
affected members in these families but not in unaffected
individuals or patients with sporadic PD. The function of
the protein encoded by this gene is unknown; however, it
is hypothesized that its mutated version clumps together
in nerve terminals causing cell death. Although this
mutation is thought to explain only a small fraction of
familial PD cases, it is hoped that the discovery can
provide clues in the other cases of PD.
Numerous other genes have been the subject of
studies attempting to link inherited factors with PD. The
cytochrome P450 family of enzymes is responsible for
detoxifying many drugs and environmental agents.95
Debrisoquine hydroxylase (CYP 2D6) is polymorphic in
nature and results in different levels of metabolism from
person to person. It is hypothesized that if exposures to
Vol 20, No 2
environmental agents play a role in PD, abnormalities in
detoxifying these agents may increase the risk of disease.
This abnormality would increase the amount of toxin
available to act on various points in the oxidative stress
model.
Earlier studies assessed subjects’ phenotypes by
orally administering debrisoquine and measuring the
amount of metabolite in urine. Subjects were labelled as
extensive metabolizers or poor metabolizers, depending
upon the percentage recovery of debrisoquine. Since no
studies of white subjects revealed a significant OR96–102
among poor metabolizers, studies focusing on
individuals’ genotype were conducted. These latter
studies involved direct analysis of the CYP 2D6 gene
and the determination of the specific variant associated
with the poor metabolizer phenotype. The studies were
very inconsistent in both their results and in the number
of variants included in the genetic analysis. Although the
results concerning the relation between PD risk and the
most common variant, CYP 2D6B,103–112 did not offer
any conclusive evidence as to whether the CYP 2D6
gene is associated with PD, it may still play a role in a
subset of individuals. Other members of the cytochrome
P450 family, such as CYP 1A2 and CYP 3A4, may also
be important in PD susceptibility.113,114
An association has been identified between the slow
acetylator genotype for N-acetyltransferase 2 and
familial PD.115 This might increase the patient’s
susceptibility to environmental toxins; however, further
study is required.
Mitochondrial gene defects
The activity of complex I, a group of proteins
involved in aerobic respiration, has been observed to be
deficient not only in the brain tissue of PD patients116–118
but also in hybrid cells,119 skeletal muscle,120,121
fibroblasts122 and, in some studies, platelets.123,124
Complex I has also been found to be inhibited by the
active metabolite of MPTP.125 Since seven of
approximately 40 subunits of complex I are encoded by
mitochondrial DNA126 and since this DNA is more easily
damaged than nuclear DNA,127 alterations in the
mitochondrial genome, whether inherited or acquired
through toxic agents, may be central to
neurodegeneration in PD.
Studies involving mitochondrial DNA mutation
analysis first reported a large deletion of genetic material
in PD patients.128 Later studies, however, downplayed
these results and suggested that this deletion was an
age-related observation, independent of PD.129–131
Although other mitochondrial gene defects have been
found in the brains of PD patients,132–135 many of these
study designs have failed to control for age. Since
mitochondrial DNA is exclusively maternally derived,
maternal inheritance of PD would be expected if
mitochondrial DNA were associated with the disease.
1999
Two studies136,137 that have specifically investigated
maternal inheritance are divided as to whether their
evidence supports the hypothesis that inheritance of an
abnormal gene is responsible for familial PD. The study
that did not support the hypothesis136 simply compared
the number of fathers and mothers of PD patients who
also had the disease, whereas the study that supported
the hypothesis137 included only those families in which
both a parent and multiple siblings had PD. The authors
of the latter study argue that simple pedigree analysis
may not be sensitive enough to detect a preponderance of
maternal inheritance.
Genetic anticipation, a phenomenon in which the
severity of disease increases in subsequent generations,
has been reported for a number of families with a history
of PD spanning multiple generations.89,138,139 This
observation is thought to be related to the expansion of
trinucleotide repeats, as is the case in diseases such as
Huntington’s disease and myotonic dystrophy.140
However, no difference in trinucleotide repeat expansion
was detected in PD patients and controls140,141 or
between generations in PD families that displayed
anticipation in age at onset.140 The pedigree analysis of
one large kindred suggested that the observation of
anticipation may be associated with an age-related
ascertainment bias.88
Since PD is thought by some researchers to be similar
to Alzheimer’s disease (AD),142,143 the apolipoprotein E
(ApoE) gene, which is linked to AD susceptibility,144–147
has been the focus of other genetic epidemiologic
studies. With the exception of one study that reported a
higher frequency of ApoE ,4 in PD patients with
dementia than in those without dementia,148 the ,4 allele
has been found to be unrelated to PD.149–156
Numerous other genetic and molecular endpoints
have been recently examined. Negative results have been
reported from studies involving superoxide dismutase,157
dopamine receptors158 and tyrosine hydroxylase,159
whereas there have been positive results for lactoferrin
receptors,160 L-cysteine,161,162 catalase activity,163 nitric
oxide163,164 and catechol-O-methyltransferase.165 One
study examined linkage for numerous genes
simultaneously in three families with autosomal
dominant inherited parkinsonism.166 Although in one
family there were slightly positive results for CYP 2D6,
there was evidence against linkage genes for glutathione
peroxidase, tyrosine hydroxylase, brain-derived
neurotrophic factor, catalase, amyloid precursor factor
and copper zinc superoxide dismutase. As with
environmental factors, there may be many genes that
play a role in PD pathogenesis. Genetic susceptibility
may limit the patient’s ability to detoxify otherwise
innocuous environmental factors and thereby lead to the
degradation of dopamine-containing neurons in the
nigrostriatal system.167
67
Monoamine oxidases
Monoamine oxidases (MAOs) are degradative
enzymes involved in the metabolism of toxins (A and B
types)32,168–170 and in the production of free radicals and
hydrogen peroxide through the breakdown of dopamine
(B type).171–173 As with the genotypic CYP 2D6 studies,
studies focusing on MAOs are very inconsistent.
Although some show a relation between PD and a
polymorphism of the gene encoding MAO type A and
not B, and others show a relation between the disease
and an MAO type B and not A polymorphism (Table
9),174–179 the evidence seems to suggest that MAO
enzyme variability may influence PD pathogenesis and
progression.
One of the first geneSenvironment interaction studies
in PD research involved an MAO-B polymorphism and
smoking.180 The study found an overall protective effect
for smoking in PD similar to the results discussed
previously. However, it further discovered that the
inverse association was only present in individuals with
a certain MAO-B variant. This breakthrough not only
adds a genetic hypothesis to the list of ideas as to how
and why smoking is protective, but it also emphasizes
the importance of both genetic and environmental factors
in PD etiology.
Diagnosis
A PD diagnosis is not necessarily clear cut, since
there is no single diagnostic test.3,181 In order for the
condition to be diagnosed, physical examination should
TABLE 9
Studies investigating the relation between
monoamine oxidase (MAO) and Parkinson’s
disease (PD)
reveal two of either tremor, rigidity or bradykinesia. All
other causes and types of parkinsonism must be
excluded. The criteria for a diagnosis of PD also include
a positive response to dopaminergic drugs such as
levodopa. PD may also be described in terms of its
severity. Hoehn and Yahr stages express the extent of an
individual’s disability on an arbitrary scale with five
levels. Stage I consists of unilateral involvement only,
usually with minimal or no functional impairment; Stage
V consists of confinement to bed or wheelchair unless
aided.182
As well, it has been reported that more than one
quarter of PD patients exhibit dementia and that some
patients with AD show signs of parkinsonism.183
According to the Merck Manual of Geriatrics, a “clinical
diagnosis is usually based on whether the motor signs
were present before or after the cognitive decline”.183
PD can only be diagnosed in an individual once
symptoms have developed; approximately 70% of
neurons in the substantia nigra have been lost when
symptoms first occur.6 This suggests an asymptomatic
period in which the disease is progressing but the
individual does not show any clinical signs. It would be
advantageous, therefore, to develop a method of
screening that would identify individuals at the earliest
stage of neurodegeneration. Intervention would then
focus on arresting the disease process rather than the
current situation of primarily treating the symptoms.
Although it is not known whether PD can be detected
before symptoms are present, numerous strategies have
been proposed.184 Studies involving positron emission
tomography (PET), movement time and the
electrophysiological characteristics of tremor show that
these methods may be useful in measuring preclinical
dysfunction.
Treatment
Study
authors
Findings
Although there is no cure for PD, both
pharmacological and surgical treatments are available.
Costa et al.179
Differences in allele frequencies of MAO-B leading to
elevated odds ratios (ORs) for the G allele in males and
females
OR was significant in females
Ho et al.176
No difference in allele frequencies for MAO-B
Hotamisligil et
al.175
Significant difference in allele frequencies for both MAO-A
and MAO-B
The main treatment for PD is pharmacological and
includes different drugs designed to either increase the
amount of dopamine in the brain or suppress the
overactive cholinergic system (anticholinergics).185–188
As dopamine cannot cross the bloodSbrain barrier, an
alternative to administering this neurotransmitter was
first introduced in the 1960s with levodopa.
Kurth et al.174
Significant difference in allele frequencies between cases
and controls for MAO-B leading to an elevated and
significant odds ratio for the G allele
No difference in allele frequencies for MAO-A
Morimoto et al.177
No difference in allele frequencies for MAO-B in Japanese
PD patients
The G allele was twice as frequent in Caucasians than in
Japanese
Nanko et al.178
No difference in allele frequencies for either MAO-A or
MAO-B
68
Chronic Diseases in Canada
Levodopa, a precursor to dopamine, has long been the
standard treatment of PD; however, it causes adverse
effects such as nausea, vomiting and orthostatic
hypotension. Although there seems to be wide agreement
that levodopa increases survival rates, there is some
debate as to when this therapy should be started.189
Levodopa therapy initially works well, but after several
Vol 20, No 2
years the majority of patients have either developed
response fluctuations (wearing off and on-off
phenomena) or dyskinesias (abnormal involuntary
movements).189 Those who assert that levodopa therapy
should be started during the early course of the disease
maintain that these motor complications reflect the
progression of the disease, whereas those who argue in
favour of delaying the drug believe that levodopa may
cause toxic effects.
Research, development and trials for more effective
drugs with fewer side effects are ongoing. Parkinsonian
symptoms can also be managed to a limited extent
through dietary modification199 and specific exercises.200
Since PD patients vary with respect to their symptoms
and disease severity, individuals will respond differently
to the same treatment. Health care professionals must
thus work alongside their patients to devise the best
possible care.
Other drugs, such as dopamine agonists,
anticholinergic agents and amantadine, have been
introduced as adjuncts to levodopa, their main function
being to minimize its adverse side effects. Other negative
effects, however, have emerged. For at least the past
decade, selegiline, a selective inhibitor of MAO-B, has
been the subject of controversy in PD treatment.190–192
There is the question as to whether this drug has a
neuroprotective effect on PD or simply a symptomatic
effect.
Prognosis and Co-morbidity
An extensive review of pharmacological treatment
approaches is beyond the scope of this Monograph
Series; thus, the reader is referred to existing in-depth
reviews.193,194
In addition to drug therapy, there are three surgical
procedures used for the treatment of PD.195,196 These are
ablative surgery, deep brain stimulation and fetal tissue
transplantation.
Ablative surgical procedures involve placing a lesion
in a circuit of either the globus pallidus (pallidotomy) or
the thalamus (thalamotomy). Since dopamine normally
modulates an inhibitory influence of the basal ganglia to
the thalamus, a dopamine deficiency would result in less
inhibition. A lesion would correct this situation in that it
would mimic dopamine in terminating nerve signals
from the globus pallidus to the thalamus. Thalamotomies
have been found to be successful for individuals with
severe tremor. Although pallidotomy is effective in
relieving bradykinesia and severe “off” motor disability,
further study is required to assess the adverse effects of
the surgery.197
Deep brain stimulation (DBS) is similar to ablative
surgery but, rather than a lesion being created, a
stimulating electrode is placed in the target. A recent
study indicates that DBS and thalamotomy are equally
successful in relieving tremor but suggests that DBS is
preferable because of the ability to alleviate side effects
and control tremor recurrence without further surgery.198
Fetal tissue transplantation has also been performed in
some PD patients. This procedure involves implanting
fetal dopamine-producing tissue into the basal ganglia in
the hope that this tissue will develop and continually
produce dopamine in the patient. Despite progress, these
procedures are still very experimental and have only
been performed in a limited number of individuals with a
small amount of time devoted to follow-up.
1999
The introduction of levodopa has increased survival
rates in PD patients. A recent study201 showed that
survival was markedly improved with its use. The
benefits were only seen, however, if levodopa therapy
was initiated in the earlier stages of disease. Although
survival rates have increased with levodopa and other
therapies, PD patients are still at increased risk of dying
as compared with individuals of similar age. A cohort
study of parkinsonian patients in England and Wales
revealed that these patients had more than a twofold risk
of dying compared with general population controls after
20 years of follow-up,202 with very little difference
between males and females. A similar study conducted
in Scotland found a similar relative risk of 2.5, but this
was only for 3.5 years of follow-up.203 Two US studies
calculated relative risks of 2.7 over a mean follow-up of
2.5 years204 and 2.0 over a mean follow-up of 9.2
years.205 A higher mortality rate was recorded for
institutionalized individuals with PD than for those
without it in one study,206 but not in another one.26
In one of the studies mentioned, the risk of death
increased with increasing number of parkinsonian signs
present.205 The presence of gait disturbance specifically
was found to be associated with increased mortality risk.
These observations must be interpreted with caution,
since some of the studies included subjects with all
forms of parkinsonism. In a study involving only
subjects with PD, the duration from disease onset to
Hoehn and Yahr stages I, II, III, IV and V were 4.0, 6.5,
7.9, 9.8 and 11.8 years, respectively.207 Patients who
noticed unilateral symptoms initially had a better
prognosis than those who first noticed bilateral
symptoms. Seventy percent of all patients noticed
unilateral symptoms initially, of which 91% spread to
the other side. Louis et al. found that the severity of
extrapyramidal signs was the single most important
indicator of increased mortality in PD patients.204
PD patients also differ from the general population in
terms of their specific cause of death. Some studies have
found that individuals with PD are more likely to die of
ischemic heart disease,202,207 cerebrovascular
disease,202,204,208 pneumonia204,208,209 and other respiratory
diseases.202 The reasons for these differences are not
known but may involve competing causes of death, a
secondary effect of treatment or common etiology.202
Cancer mortality has been found by some to be
69
significantly lower in PD patients than in an age- and
sex-matched population.204,208 However, this is only true
for cancers that are thought to be related to smoking and
is explained by the fact that PD patients are less likely to
smoke.210
Depression
Depression has been found to be more prevalent in
individuals suffering from physical illnesses such as
stroke, cancer and endocrine and metabolic disorders.211
The few recent studies that have examined the relation
between depression and PD have been very inconsistent.
This variability is largely due to the different criteria
used to measure depression. Recent studies, however,
have agreed that depression is more common in PD
patients with dementia than patients without
dementia.212–215 Depression in PD patients was also
found to be related to thought disorders215 and autonomic
failure.216 Inconsistencies have been noted for the
relation between depression and age, disease course and
impairment in the activities of daily living.214,215,217,218
References
1.
2.
3.
4.
5.
6.
7.
8.
Conclusions
Implications for an Aging Population
Since the proportion of both Canada’s and the world’s
population that is over 65 years of age will increase
dramatically over the next three decades, the number of
individuals with PD is expected to increase
correspondingly. In fact, the percentage of Canada’s
population over the age of 65 is expected to increase
from 11.6% in 1991 to 23.6% by the year 2016.219 This
translates into an 87% increase in the number of
individuals requiring medical care on an in-patient basis
and a 92% increase in the number of individuals dying
from PD. The change will be greatest in the oldest age
groups, in which the number of individuals affected with
PD is expected to more than double.
Future Research
Although the past few years of PD research have been
very productive, many questions remain.220 The
identification of biological markers for PD will play an
integral part in future research. This will enable
researchers and physicians to diagnose the disease more
accurately, develop treatments that slow or arrest disease
progression once it has started and initiate steps towards
preventing the disease. It would also allow individuals
known to be at higher risk to be monitored before
symptoms appear. Genetic research may lead to the
identification of other genes that predispose individuals
to PD, and further investigation of the role of
environmental risk factors could supply important
information for the development of prevention strategies.
An understanding of the interplay between
environmental and genetic factors could provide the key
to future advances in PD research.
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Vol 20, No 2
Development of Record Linkage of Hospital Discharge
Data for the Study of Neonatal Readmission
Shiliang Liu and Shi Wu Wen
Abstract
Computerized record linkage has been used increasingly in epidemiologic studies. We
developed a multi-stage, deterministic matching algorithm using various combinations of key
variables. Then, from the records for March 1, 1993, to March 31, 1996, contained in the
discharge abstract database of the Canadian Institute for Health Information (CIHI), we
examined the relation between length of hospital stay at birth and neonatal readmission. A
combined use of province/territory of occurrence, 6-digit postal code of residence, date of
birth and sex (step 1) matched 88.5% of 26,629 eligible neonatal readmission records with
their birth records. Additional use of institution code and chart number or health card number
combined with date of birth and sex (step 2 and step 3) increased the matching rate to 93.0%.
Compared with the gold standard, step 1 correctly matched 94.4% of the records. We
conclude that this deterministic matching algorithm is a feasible and convenient approach to
data linkage for the study of neonatal readmission. The linkage strategy may also be helpful in
epidemiologic studies of other short-term events.
Key words: epidemiologic method; hospital discharge abstract; medical record linkage;
neonatal readmission
Introduction
Studies of existing databases are attractive to
epidemiologists and other health researchers because
they can be done efficiently at the level of large
populations. For example, it is possible to examine the
relation between birth weight, gestational age, maternal
age and infant mortality or morbidity at the country level
by analyzing existing data, as the information is
routinely recorded in vital and hospital statistics.
However, the lack of comprehensive information in a
single database often impedes researchers in this effort.
In recent years, the development of computerized record
linkage has made it possible to overcome such obstacles
in existing database studies.1S17
Record linkage methods can be summarized into three
broad categories: manual, deterministic and probabilistic.
Manual matching is the oldest, most time-consuming and
most costly method, but remains the standard. However,
it is not a feasible option when large databases are
involved. Probabilistic linkage is used to identify and
link records from one data set to corresponding records
in another data set (or two records from different
locations in a single data set) on the basis of a calculated
statistical probability for a set of relevant variables (e.g.
name, sex, date of birth). Deterministic linkage matches
records from two data sets (or two records from different
locations in a single data set) using a unique variable
(e.g. social insurance number or hospital chart number)
or by full agreement of a set of common variables (e.g.
name, sex, birth date).
Probabilistic linkage is considered the preferred
method, because the calculation of the probability can be
refined in various respects to accommodate weights
associated with identifier values and coding errors, thus
maximizing the available information in the data.1S3,16,17
However, the probabilistic linkage requires detailed prior
knowledge about various measures of the relative
importance of specific identifier values—for example,
frequency—in both files that are to be linked.
Investigators often do not have this degree of prior
knowledge.6
This paper aims to illustrate the use of deterministic
linkage of hospital discharge records in the hospital
Author References
Shiliang Liu and Shi Wu Wen, Bureau of Reproductive and Child Health, Laboratory Centre for Disease Control, Health Canada, Tunney’s Pasture,
Address Locator: 0601E2, Ottawa, Ontario K1A 0L2
1999
77
discharge database of the Canadian Institute for Health
Information (CIHI), taking neonatal readmission as an
example. One of our previous studies revealed a
substantial recent reduction in length of newborn
hospital stay at birth.18 We hypothesized that this
reduction might increase rates of neonatal readmission.
To allow an examination of the relation between length
of newborn hospital stay at birth and subsequent
neonatal readmission, a linkage of readmission record
with the infant’s own birth record is required.
Methods
Three years of CIHI data (fiscal years 1993/94 to
1995/96) were used. Data for Nova Scotia, Quebec and
Manitoba were excluded because CIHI collected only a
small proportion of hospital discharge records in these
provinces.19 Live infants were identified by a field of
“age unit” with a code of “NB.” Infants weighing less
than 1500 g, those discharged from hospital after 21 days
from birth and those who subsequently died in their
hospital of birth were excluded. A neonatal readmission
was defined as admission of an infant to any hospital,
within 28 days of birth. Infants who were transferred
from another institution were not included as
readmission cases. Multiple births were excluded from
both birth and readmission records because
non-identifiable variables were shared among them.
Both birth and readmission records have information
on province/territory and institution of occurrence,
institution chart number, date of birth, sex, provincial
health card number, 6-digit postal code, admission date,
discharge date and diagnostic codes. Institution code,
institution chart number and provincial health card
number are scrambled for confidentiality considerations
(Table 1).
Theoretically, the health card number and/or
institution chart number, although scrambled, can be
used as a unique variable for record linkage because the
same number is used for each individual once it has been
assigned by the provincial/territorial authority or
hospital. However, because of delay in obtaining the
health card number, infants are usually assigned their
mother’s number or that field is left blank at birth. We
were concerned that using the health card number alone
might lead to confusion or error if infants were
subsequently given their own number or shared a
number with their siblings. The institution chart number
is effective only when an infant is readmitted to the
hospital where he or she was born; only a small
proportion of cases were readmitted to the hospital of
their birth, however.
Accordingly, we considered it appropriate to use a set
of variables for multi-stage deterministic linkage. Based
on our assessment of the availability and appropriateness
of the variables on CIHI discharge records, a computer
matching algorithm was designed. As described in
Figure 1, records of birth and readmission were matched
78
Chronic Diseases in Canada
TABLE 1
Availability of proposed matching variables for
record linkage in birth file and in neonatal
readmission file
Variable
Birth file
Readmission file
Number of records
788,480
27,405
Province (%)
100.0
100.0
Institute number (%)
100.0
100.0
Chart number (%)a
97.4
98.2
Health card number (%)b
86.8
80.6
Postal code (%)c
97.9
98.0
Residence code (%)
70.1
71.0
Date of birth (%)
100.0
100.0
Sex (%)
100.0
100.0
Admission date (%)
100.0
100.0
Discharge date (%)
100.0
100.0
a
Different institutions have different chart number series. Only an infant who is
readmitted to the same hospital of birth is assigned the identical chart
number.
b
It was found that a majority of infants were assigned their mothers’ health card
number at birth or at readmission.
c
About 1% of records showed no information on postal code; another 1%
contained incomplete 6-digit postal codes in both files.
first by full agreement of province/territory of
occurrence, 6-digit postal code of residence, date of birth
and sex (step 1); second, by full agreement of institution
code, institution chart number, sex and date of birth
(step 2); and third, by full agreement of provincial/
territorial health card number, sex and date of birth
(step 3); finally, matching was supplemented by a logic
check of the matched cases (step 4). The logic check
involved determining whether there were conflicts or
contradictions between birth date, discharge date,
readmission date and age at readmission.
To evaluate the accuracy of the record linkage carried
out in step 1, on which the majority of the successful
matches were based, we created a linked file by using
step 2 alone to identify the infants who were readmitted
to the hospital of birth. We considered this linked file as
the gold standard, because the institution chart number is
unique in these records. We then separated the linked
birth and readmission records, and performed step 1 to
link them again in order to assess its matching accuracy
as compared with that of the gold standard.
Finally, we assessed the potential bias caused by
exclusions and unsuccessful linkage by comparing the
distributions of variables of interest, such as birth
weight, length of hospital stay and main diagnostic
categories for readmission, between the linked and
unlinked cases. In this comparison, the unlinked cases
included those who were excluded according to the
selection criteria before the linkage procedure was
performed. SAS software for Unix, version 6.12 (SAS
Vol 20, No 2
FIGURE 1
Matching algorithm for record linkage
of hospital discharge data
in hospital or who were part of multiple births, we found
798,840 live birth records that met the inclusion criteria.
During the corresponding period, a total of 27,405
infants in the same nine Canadian provinces and
territories were readmitted to hospitals within 28 days of
birth. According to the selection criteria, 26,629 of these
readmissions were eligible to be linked with birth
records.
Step 1 successfully matched 23,571 readmitted infants
(after excluding 26 duplicates) to their birth records,
accounting for 88.5% of the 26,629 eligible readmission
cases. Implementation of steps 2 and 3 increased the
successful matches to 24,766 readmission cases,
representing 93.0% of eligible readmission cases, after
two pairs were excluded by step 4 (logic check). Details
of the matching process are given in Figure 1.
Among the 7430 cases in the linked file used as the
gold standard, 7023 (94.5%) cases were successfully
matched by implementation of step 1 as described in
Figure 1. Of these 7023 cases, 2 cases were falsely
matched and 7 were duplicates, as a result of their
non-identical matching variables. Therefore, the correct
matching rate was 94.4% using step 1, i.e. full agreement
of province of occurrence, 6-digit postal code of
residence, sex and date of birth.
Comparison of linked and unlinked cases showed that
they were quite similar in main characteristics and
diagnoses of interest (Table 2). However, statistically
significant higher proportions of infants of low birth
weight (6.4% versus 5.6%) and readmissions with a
diagnosis of jaundice (40.9% versus 38.6%) were
observed in unlinked cases. There was also an increase
in the rate of successful record linkage from fiscal year
1993/94 to 1995/96 (Table 2).
Discussion
Probabilistic matching is a recommended strategy for
computerized record linkage. It is considered the
preferred method because the calculation of the
probability can be refined in various respects to
accommodate weights associated with identifier values
and coding errors, thus maximizing the available
information in the data.1S3,16,17
Institute Inc., Cary, North Carolina), was used in all data
abstraction and linkage processing.
Results
A total of 817,351 live infants were born in hospitals
in the nine Canadian provinces and territories studied
and were recorded by CIHI during the period of March
1, 1993, to March 31, 1996. After excluding infants who
weighed less than 1500 g, who were discharged from
hospital after 21 days from birth, who subsequently died
1999
If there is a common unique identifier (e.g. social
insurance number) in both files to be linked, and if the
common unique identifier is quite accurately recorded in
the data, deterministic linkage can be performed
conveniently by using routine statistical software such as
SAS. However, such a common unique identifier is often
not available. For example, social insurance numbers or
other personal identifiers are often issued to adults only,
so that they cannot be used in studies involving infants
and children. For confidentiality considerations, the data
collector is often prohibited from releasing the subject’s
name. Even if the subject’s name can be released to
investigators, spelling mistakes in names are frequent.15
79
TABLE 2
Comparison of main characteristics of linked and unlinked
cases in a study of neonatal readmission
Characteristic
a
Linked cases
Unlinked cases
24,766
2,639
% of fiscal year 1993/94
30.8
35.2
% of fiscal year 1994/95
33.4
33.2
NS
% of fiscal year 1995/96
35.8
31.6
<0.01
% of males
57.0
56.3
NS
Number
% of birth weight <2500 g
p value
<0.01
5.6
6.4
<0.01
Mean age at readmission (days)
10.8
10.7
NS
% of length of stay <2 days at birth
25.6
25.8
NS
% of infants with jaundice
40.9
38.6
<0.05
% of infants with dehydration
5.9
6.1
NS
% of infants with inadequate weight gain
2.8
2.4
NS
% of infants with feeding problems
9.8
10.2
NS
% of infants with sepsis
5.4
5.3
NS
a
The number includes the cases that were excluded prior to linkage procedure by subject selection criteria.
NS = Not significant
Postal code is a well-developed system of Canada
Post Corporation. This information is often recorded
completely, and the chance of a mistake is relatively low,
as the code tends to be shorter and simpler than name
and address. In addition, because it does not reveal an
individual’s identity, it can be fully released to
investigators without confidentiality concerns. We
performed a frequency procedure on our raw data, and
found that the chances of two individuals sharing the
same sex, date of birth and 6-digit postal code were very
low (data not shown). With combined use of sex, date of
birth and other information, this variable can play a key
role in identifying the same individual. This procedure
(i.e. step 1) accounted for the majority of the linked
records (88.5%) in our study of neonatal readmission; as
well, it was quite accurate (94.4% as compared with the
gold standard). In Canada, a small proportion of births
occur outside of hospitals. If we had access to data on
out-of-hospital births, the matching rate would be even
higher.
As with other linkage methods, the success of
deterministic linkage depends largely on the
completeness and accuracy of the information in the files
to be linked and an appropriate combination of matching
variables. In our linkage procedures, failure in matching
was largely caused by missing or incomplete information
on the variables used, such as postal code. However, as
suggested by the increasingly successful matching rates
from year 1993/94 to 1995/96 (Table 2), the quality of
CIHI hospital discharge data is improving, and this
provides promise for future studies using deterministic
linkage.
80
Chronic Diseases in Canada
When failure in linkage occurs, it is important to
assess its potential impact on the study results. One
consequence is that the sample size available for analysis
will be reduced. However, because sample size is usually
not an issue in existing database studies, the real concern
of incomplete record linkage is the potential bias
introduced by unsuccessful linkage. Our comparison of
linked and unlinked cases showed no substantial
differences in main characteristics and diagnostic
categories of interest (despite statistically significant
differences in low birth weight and jaundice rates),
suggesting that no major bias was introduced by this
record linkage.
One limitation of record linkage using postal code as
a key matching variable should be emphasized. In
modern society, people relocate quite frequently. As a
result, deterministic record linkage involving postal code
may be less reliable in studies of long-term events. In our
case, the chance of relocation within 28 days after birth
was low, unless the patients gave different addresses at
different hospital admissions (e.g. gave parents’ address
at birth but grandparents’ address at readmission). In
addition, relying on full agreement of a set of matching
variables often restricts some potential matching pairs or
reduces the sensitivity.
The deterministic matching algorithm provided a
feasible and convenient approach to data linkage for our
study of neonatal readmission. Although it was
developed for a specific purpose, it may also be used for
epidemiologic studies of other short-term events, such as
epidemic outbreak, rehospitalization, adverse drug or
vaccine reactions and familial aggregation of disease or
Vol 20, No 2
risk factor. For example, with some modifications of the
linkage program, it may be used to study maternal
readmission or the relation between maternal
characteristics and infant outcomes.
10.
Acknowledgements
This study was carried out under the auspices of the Canadian
Perinatal Surveillance System. The authors thank Dr Catherine
McCourt for her comments on the manuscript.
11.
References
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Miller AB, Howe GR, Sherman GJ. Mortality from breast
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Herderson J, Goldacre MJ, Graveney MJ, Simmons HM.
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Goldberg MS, Carpenter M, Thériault G, Fair M. The
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Howe GR. Lung cancer mortality between 1950 and 1987
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Howe GR, McLaughlin J. Breast cancer mortality
between 1950 and 1987 after exposure to fractionated
moderate-dose-rate ionizing radiation in the Canadian
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Herrchen B, Gould JB, Nesbitt TS. Vital statistics linked
birth/infant death and hospital discharge record linkage
for epidemiological studies. Comput Biomed Res
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Adams MM, Wilson HG, Casto DL, Berg CJ, McDermott
JM, Gaudino JA, McCarthy BJ. Constructing
reproductive histories by linking vital records. Am J
Epidemiol 1997;145:339–48.
Waien SA. Linking large administrative databases: a
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Howe GR. Use of computerized record linkage in cohort
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Wen SW, Liu S, Fowler D. Trends and variations in
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Wen SW, Liu S, Marcoux S, Fowler D. Uses and
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1997;18(3):113–9. O
81
Rate and Cost of Hospitalizations for Asthma in Quebec:
An Analysis of 1988/89, 1989/90 and 1994/95 Data
Claudine Laurier, Wendy Kennedy, Jean-Luc Malo, Michèle Paré, Daniel Labbé, André Archambault
and André-Pierre Contandriopoulos
Abstract
The objectives of this study were to evaluate recent trends in the frequency and length of stay
of hospitalization for asthma in the province of Quebec and to estimate the costs of asthma
hospitalizations. Data were extracted for persons hospitalized for 30 days or less with a
primary diagnosis of asthma in all Quebec short-stay hospitals during the years 1988/89,
1989/90 and 1994/95. There were 1.76 asthma hospitalizations per 1000 persons in Quebec
in 1988/89, down to 1.44 in 1989/90 and up again to 1.75 in 1994/95. There was a small
decrease in mean length of stay when the three data years were compared. In all three years,
the rate of hospitalization was particularly high among young boys. In 1994/95, more
hospitalizations occurred during the fall months. We estimated the total cost for asthma
hospitalization that year to be $18 to $21 million.
Key words: asthma; costs; hospitalizations; Quebec
Introduction
Asthma is a common disease of both adults and
children, causing breathing impairment with
consequences dependent on the severity of the disorder.
Studies have estimated the population prevalence as 3%1
and 7%2 in the adult population and over 10% in
children.3 According to the 1987 Santé Québec survey,
the prevalence of asthma, bronchitis and emphysema (as
one comprehensive category) was 3.9% in the general
Quebec population.4 Santé Québec has published the
prevalence by sex (4.1% among women and 3.9%
among men) and by age group for some groups: 4.1%
among those aged less than 15 years, 3.2% among those
15S24, 4.4% among those 45S64 and 7.1% among those
65 and older.4
Rates of Quebec hospitalizations for asthma increased
during the decade from the early 1970s to the early
1980s, by 79% to 1.56 per 1000 among males and by
58% to 1.33 per 1000 among females.5 The early 1980s
may have seen a decrease: in 1984/85 there were 9080
hospitalizations for asthma in Quebec, down from
11,726 in 1980/81.6 Over the entire 1980s there was an
average increase in the annual rate of hospitalizations of
3.3%.7 In Canada, the number of hospitalizations
increased at an average rate of 1.8% per year from 1981
to 1989, and the mean length of stay decreased from 6 to
4.7 days over the same period.8 Hospitalization rates
differed across age groups—in 1988 the estimated rate
was 5.82 per 1000 among those under 15 years old and
0.66 among those aged 15 to 34.8 Asthma hospitalization
was subject to a substantial seasonal influence, the rate
peaking in autumn.9, 10
The preceding data describe the situation in the
1980s. With the increase in use of inhaled corticosteroids
(from 6 per 1000 Saskatchewan inhabitants in 1989 to 20
per 1000 in 199311), reflecting the dissemination of
guidelines and recommendations for treatment of
asthma, there could be a decrease in the need for
Author References
Claudine Laurier, Faculté de pharmacie and Groupe de recherche interdisciplinaire en santé, Université de Montréal, Montreal, Quebec
Wendy Kennedy and André-Pierre Contandriopoulos, Administration de Santé and Groupe de recherche interdisciplinaire en santé, Université de
Montréal, Montreal, Quebec
Jean-Luc Malo, Faculté de médecine, Hôpital du Sacré-Cœur, Université de Montréal, Montreal, Quebec
Michèle Paré, Groupe de recherche interdisciplinaire en santé, Université de Montréal, Montreal, Quebec
Daniel Labbé, Direction générale de la planification et de l’évaluation, Ministère de la Santé et des Services sociaux du Québec, Quebec
André Archambault, Faculté de pharmacie, Université de Montréal, Montreal, Quebec
Correspondence: Claudine Laurier, Faculté de pharmacie, Université de Montréal, C.P. 6128, succursale Centreville, Montréal (Québec) H3C 3J7
82
Chronic Diseases in Canada
Vol 20, No 2
hospitalization due to asthma. Such a decrease was seen
recently in Sweden.12 Offsetting this expected trend is
the increase in the prevalence of asthma that is generally
agreed to be taking place.13
Inasmuch as the cost of asthma treatment is important,
relatively up-to-date estimates of the cost of
hospitalization for asthma—a major component of
overall costs—are of interest. The direct costs of asthma
in Canada for 1990 were estimated to be $306 million;
hospital in-patient care was an estimated $84.4 million,
excluding drugs.14
To explore the hypothesized changes in rates of
hospitalization and to estimate the cost of this area of
asthma treatment, we examined rates of hospitalization
with a primary diagnosis of asthma for the Quebec
population during three one-year periods: 1988/89,
1989/90 and 1994/95. Rates and average lengths of stay
of such hospitalizations were established, and those for
1994/95 were examined according to age, sex and month
of admission. The costs associated with the 1994/95
asthma hospitalizations were then estimated.
Methods
This study used the MED-ECHO database for fiscal
years 1988/89, 1989/90 and 1994/95 (i.e. from April 1
until March 31 for each year). MED-ECHO is an
electronic database that contains detailed hospitalization
summaries. The database includes information on the
principal diagnosis and up to 16 associated diagnoses,
dates of admission and separation, length of stay, ward,
major procedures undergone as well as the age and sex
of the patient. All hospitalizations with the principal
diagnosis of asthma (ICD-9 codes 493.0 to 493.9) were
retrieved for the relevant years. As MED-ECHO annual
files are organized according to date of separation, the
hospitalizations analyzed in this study included those in
which the patient was admitted during the previous
period but discharged in the period of study. On the other
hand, hospitalizations for patients admitted during the
period of study but discharged in the subsequent period
were not included. Less than 1% of retrieved
hospitalizations were in long-stay hospitals or were for a
period of greater than 30 days. These long-stay
hospitalizations were excluded from the analysis, as they
were likely to be related to conditions other than simple
asthma.
Population data represented the population eligible for
the Health Insurance Program on July 1st of the
respective year (1989, 1990 and 1995) and was lower by
about 86,000 than the total estimated population of
roughly 7 million.15 It should be noted that MED-ECHO
reports the number of hospitalizations and not the
number of people hospitalized in any given period. The
rates presented here must be interpreted accordingly.
Rate of hospitalization per 1000 and length of stay
were estimated by age group and sex for the three
1999
one-year periods. Age (in years) was grouped as follows:
less than 1, 1S4, 5S9, 10S14, 15S19, 20S39, 40S64,
65S74, and 75 and older. The 1994/95 hospitalization
rates were also analyzed by month of admission.
Differences in the average length of stay among patients
according to the month of admission, age group and sex
were tested for significance using parametric or
non-parametric tests (t-test, analysis of variance
[ANOVA] or Kruskall-Wallis). A two-way ANOVA was
performed for age and sex.
The cost of asthma hospitalization was estimated by
means of two methods. The first used the 1994/95
financial data for all short-stay hospitals (excluding
psychiatric hospitals). A per diem specific to the hospital
ward was calculated, which included nursing care,
pharmacy, laboratory, “hotel,” administration and
maintenance costs; it excluded capital cost investment
and physician reimbursement. The recorded length of
stay was multiplied by the per diem.
The second approach used an index reflecting the
relative use of resources (NIRRU) for each
hospitalization classified according to its All-Patient
Refined Diagnostic-Related Group (APR-DRG).16 These
APR-DRGs constitute 1530 groups of clinically
homogeneous patients requiring an equivalent level of
resources. They are based on diagnosis, severity and
probability of poor outcome. Among other changes to
the previous versions of the DRG classifications, the
APR-DRG system has incorporated certain specific
pediatric DRGs. For each APR-DRG, a NIRRU was
created.
The NIRRU was based on costs per APR-DRG for
typical patients established in Maryland in 1994. The
cost for each APR-DRG was divided by the mean cost
for all hospitalizations to obtain a relative weight, where
1.00 corresponded to the mean. This index was then
adjusted to take into account the differences in lengths of
stay between Maryland and Quebec (see Appendix). A
NIRRU was calculated for each asthma hospitalization
and was applied to the average cost of hospitalizations in
the province of Quebec for 1994 to establish a cost per
asthma hospitalization.
Total and average costs for all asthma hospitalizations
were estimated using both methods, and 95% confidence
intervals were calculated for average costs per
NIRRU-adjusted asthma hospitalization.
Results
Hospitalization Rates for Asthma
From 1989 to 1995, the population of Quebec eligible
for health benefits increased by 4.1%, from 6.91 million
to 7.19 million. The rate of the total number of
hospitalizations less than 30 days (for any reason) at the
end of that period was 14.8% higher than at the
beginning, increasing from 142 per 1000 in 1988/89 to
83
December) and lower in the summer months (from May
to August), particularly in the holiday months of July
and August (Figure 2).
163 per 1000 in 1994/95. The proportion of Quebec
hospitalizations associated with asthma as the principal
diagnosis was 1.1% in 1994/95, slightly lower than in
1988/89 (1.2%). The rate of hospitalization with asthma
as the principal diagnosis was 1.76 per 1000 in 1988/89,
fell to 1.44 in 1989/90 and rose to 1.75 in 1994/95
(Table 1).
Length of Hospital Stay
The average length of stay for asthma hospitalizations
decreased from one study period to another. Mean
lengths of stay were significantly longer for women than
men in all three periods. Mean length of stay for men
decreased from 1988/89 to 1994/95. It stayed roughly the
same for women in 1988/89 and 1989/90, but was lower
in 1994/95.
Rates were estimated at 1.85 per 1000 among men
and 1.67 per 1000 among women in 1988/89, falling to
1.79 and 1.71 per 1000, respectively, in 1994/95
(Table 1).
The hospitalization rate for children
less than 1 year old in 1994/95 was twice
as high as in 1988/89 or 1989/90. In the
1S4 age group, it remained relatively the
same but slightly lower than the 1988/89
figure.
FIGURE 1
Rates of Quebec hospitalizations with asthma as principal
diagnosis (LOS #30 days), by age and sex, 1994/95
Figure 1 shows that hospitalization
rates for asthma in 1994/95 were highest
among male infants under the age of 1
year, and high among boys aged 1S4. As
well, boys under 5 were hospitalized
at a rate roughly twice that of their
female counterparts. In the age groups
of 10 years and older, however, the rates
among women were higher, and females
aged 20S39 were hospitalized at roughly
twice the rate of their male counterparts.
For 1994/95, the hospitalization rates
for asthma were higher in the autumn
and winter months (from September to
TABLE 1
Quebec hospitalizations (number and rate per 1000 population) with asthma as principal diagnosis
(LOS #30 days) and mean length of stay (LOS), 1988/89, 1989/90 and 1994/95
Sex and
age
group
1988/89
n
Rate
Mean
LOS
1989/90
LOS 95%
CI
n
Rate
Mean
LOS
1994/95
LOS 95%
CI
n
Rate
Mean
LOS
LOS 95%
CI
TOTAL
12,166
1.76
4.27
4.20–4.34
10,217
1.44
4.13
4.05–4.21
12,604
1.75
3.75
3.69–3.81
Males
6,328
1.85
3.75
3.67–3.83
5,179
1.48
3.47
3.38–3.56
6,335
1.79
3.19
3.12–3.26
Females
5,838
1.67
4.83
4.72–4.94
5,038
1.41
4.81
4.68–4.94
6,269
1.71
4.31
4.21–4.41
<1
584
6.20
4.41
4.12–4.70
521
5.48
3.98
3.70–4.26
1,121
12.30
3.54
3.39–3.69
1–4
3,475
12.24
3.05
2.97–3.13
4,283
12.20
2.63
2.57–2.69
4,315
11.08
2.38
2.33–2.43
5–9
1,248
4.01
3.03
2.91–3.15
1,917
4.10
2.63
2.55–2.71
1,341
3.01
2.33
2.25–2.41
10–14
799
2.27
3.30
3.14–3.46
1,062
2.19
2.94
2.83–3.05
859
1.79
2.83
2.70–2.96
15–19
461
1.06
3.51
3.28–3.74
494
1.08
3.18
2.95–3.41
582
1.22
2.98
2.78–3.18
1,197
0.61
4.43
4.23–4.63
1,310
0.57
4.01
3.83–4.19
20–39
1,098
0.48
4.37
4.17–4.57
40–64
1,507
0.88
6.77
6.52–7.02
1,658
0.83
6.60
6.35–6.85
1,710
0.79
5.72
5.50–5.94
65–74
617
1.42
8.06
7.60–8.52
617
1.34
8.44
7.98–8.90
739
1.45
7.36
6.96–7.76
84
Chronic Diseases in Canada
Vol 20, No 2
Mean length of stay also varied significantly with
respect to age group (p < 0.01, chi-squared = 2874,
Kruskall-Wallis). Data for all three periods showed a
consistent pattern of relatively long duration for the
youngest group (less than 1 year), declining to the
shortest mean duration in the 5S9 age group, then
increasing up to the longest stay in the 75+ age group
(Table 1).
The two-way ANOVA revealed a significant
interaction of age and sex on mean length of stay in
1994/95 (p < 0.001, F = 21). Mean lengths of stay and
95% confidence intervals (CIs) for age group and sex
appear in Table 2. For boys and girls under age 5, mean
lengths of stay appeared similar. Differences between
men and women were significant in three age groups
(5S9, 10S14 and 40S64).
The average length of stay in 1994/95 also varied
significantly according to month of admission (Figure 2),
being shorter for those admitted during the fall months
(p < 0.01, chi-squared = 221, Kruskall-Wallis).
Cost of Hospitalizations for Asthma
The results of the cost estimation for asthma
hospitalizations in Quebec for the year 1994/95 are
presented in Table 3. For the first method of calculation,
using the per diem for short-stay hospitals ($379),17 the
average 1994/95 cost per stay was approximately $1400
and the total cost was $17.9 million.
FIGURE 2
Quebec hospitalization rates with asthma as principal diagnosis
(LOS #30 days) and average length of stay, by month, 1994/95
A NIRRU was estimated for all
1994/95 short-stay asthma
hospitalizations except 54 (0.4%). The
average NIRRU for all the asthma
hospitalizations was 0.6 (±0.003), and
this varied according to the ward in
which the patient was treated (Table 3).
The average cost per stay was $1676
(95% CI = 1661S1692). Of the total cost
of $21.0 million, the greatest proportion
was accounted for by the Pediatrics ward
($9.9 million). The NIRRU-adjusted cost
per stay was lowest in Allergy ($1397)
and Pediatrics ($1,413) wards, where the
average lengths of stay were shortest,
and highest in Internal Medicine ($2150)
and Other ($2348) wards, where the
average lengths of stay were highest.
Discussion
TABLE 2
Quebec hospitalizations with asthma as principal
diagnosis (LOS #30 days) and mean length of stay (LOS),
by age and sex, 1994/95
Age
group
Hospital separations: MALES
n
Mean
LOS
(days)
LOS 95%
CI
Hospital separations: FEMALES
n
Mean
LOS
(days)
LOS 95%
CI
<1
783
3.57
3.38–3.7
4
338
3.49
3.21–3.7
7
1–4
2,824
2.40
2.33–2.4
7
1,491
2.36
2.28–2.4
4
5–9
766
2.23
2.13–2.3
3
575
2.46
2.34–2.5
8
10–14
410
2.55
2.39–2.7
1
449
3.09
2.88–3.2
9
15–19
188
2.72
2.43–3.0
1
394
3.10
2.83–3.3
7
20–39
443
3.88
3.59–4.1
6
867
4.08
3.85–4.3
1
1999
Earlier studies found an increase in
the hospitalization rate for asthma in
Quebec from the early 1970s to the early
1980s.5 The hospitalization increases
during that period for all of Canada were
greatest in persons under the age of 15.5
From 1981 to 1988, the overall Canadian
age-standardized rate of hospitalizations
rose by roughly 40%. Although the rate
in Quebec was lower than the national
average, the increase was relatively high,
at 70% and 77% for those under 15 and
ages 15S35, respectively.8
Our data show that asthma
hospitalization rates were lower in
1989/90 than in 1988/89. As well, mean
length of stay was shorter in 1989/90
than in 1988/89. In 1994/95, a return to
the 1988/89 level of asthma
hospitalization rate was seen; however,
the mean length of stay was shorter than
in the preceding period. The decrease in
85
TABLE 3
Costs of Quebec hospitalizations with asthma as principal diagnosis (LOS #30 days),
by hospital ward and calculation method, 1994/95
Hospital
ward
Hospital separations
Costs: METHOD A
a
n
% of total
Mean LOS
TOTAL DAYS
Cost per day
Cost per stay
Allergy
439
3.5
2.19
961
$379
$831
$364,817
Internal
medicine
587
4.7
6.05
3,551
$379
$2,296
$1,347,595
Pneumology
1,862
14.8
5.70
10,613
$379
$2,163
$4,027,361
Pediatrics
7,039
55.8
2.60
18,301
$379
$987
$6,944,649
Medicine
2,475
19.6
5.08
12,573
$379
$1,928
$4,770,951
202
1.6
6.19
1,250
$379
$2,349
$474,469
12,604
100.0
3.75
47,251
$379
$1,423
$17,929,842
Other
TOTAL
Hospital
ward
TOTAL COST
Costs: METHOD Ba
Hospital separations
b
% of total
Allergy
439
3.5
0.497
0.484–0.511
$1,397
$1,358–1,436
$613,182
Internal
medicine
580
4.6
0.766
0.729–0.803
$2,150
$2,047–2,253
$1,247,046
Pneumology
1,849
14.7
0.725
0.709–0.741
$2,037
$1,992–2,081
$3,765,525
Pediatrics
7,037
56.1
0.503
0.499–0.508
$1,413
$1,400–1,425
$9,941,170
$5,001,394
Medicine
Other
TOTAL
n
Mean NIRRU
Cost per stay
95% CI
TOTAL COST
2,445
19.5
0.729
0.712–0.745
$2,046
$2,000–2,091
200
1.6
0.836
0.763–0.909
$2,348
$2,143–2,553
$469,552
12,550
100.0
0.597
0.592–0.603
$1,676
$1,661–1,692
$21,037,893
mean length of stay is consistent with that reported
elsewhere for Canada.18
Our analysis has confirmed the differences according
to age and sex that have been found in previous
studies.1,8 Longer mean hospital stays were associated
with individuals less than 1 year old and those over 75.
There was an interaction of the variables of age and sex
with respect to length of hospital stay: mean stays were
relatively similar until the age of 5, and after that the
length of stay was often longer for girls and women.
This could be explained by an association between
average length of hospital stay and the presence of
comorbid conditions, more frequent in older women.
Indeed, hospitalizations with a secondary diagnosis
were, on average, of longer duration (4.3 days, 95%
CI = 4.17–4.34) than those without (2.6 days, 95%
CI = 2.52–2.65).
As documented previously,9,10 most asthma
hospitalizations occurred in the fall and the fewest, in the
summer. The reduced hospitalization rate in the summer
was possibly partially due to holidays and hospital bed
closures associated with health care staff reductions. The
relatively high numbers in the fall may be associated
with increased viral infections occurring near the
beginning of the school year19 and ragweed allergy
season (from mid-August until the end of September),20
86
NIRRU 95% CI
Chronic Diseases in Canada
the asthmatic individual tolerating the reaction for a
certain length of time until hospital admission was
necessary. An additional cause could be the increased
time spent indoors during these fall months, with
increased exposure to indoor allergens such as mites and
domestic animals.
This seasonal peak does not appear to be an artifact of
including patients with chronic obstructive pulmonary
disease misdiagnosed as asthma (whose problems may
increase in the fall months) because the increase, when
analyzed by age group, seemed to be accounted for
mostly by children under the age of 10. There was a
small rise in the number of admissions in the
40S64-year-old group in September and October, but not
in those aged 65 and over (data not shown).
Length of hospital stay was also associated with
month of admission, in that the shortest stay was
associated with fall admission and the longest with
winter admission. Although at first glance this could
reflect the availability of hospital beds, since the length
of stay appeared to be inversely related to the number of
admissions, it was probably more likely due to the
increased numbers of hospitalizations of young children
during the fall, with their concomitant shorter average
stays.
Vol 20, No 2
The NIRRU-adjusted cost estimate was higher than
that calculated from the average hospitalization cost.
Most likely this was due to the ability of the NIRRU to
account for the higher cost of days at the beginning of a
hospitalization. Asthma hospitalizations are shorter, on
average, than overall hospitalizations in Quebec, by
almost 50%. Economic evaluations that apply average
costs per hospital day could be underestimating the true
cost.
The NIRRU index was based upon costs in Maryland,
US, and this could limit its usefulness. It is assumed that
although the absolute costs per APR-DRG could differ
between Maryland and Quebec, the relative costs of one
group compared with another should be the same. The
adjustments for the Quebec situation did take into
consideration differences in the average length of stay
between Maryland and Quebec (Quebec hospital stays
are normally longer) and the fact that the end of a
hospital stay was less resource-intensive than the
beginning. There are limitations, however, in using an
index based on a different system: we cannot account for
structural differences in costs, such as for nursing care
and equipment, nor do we have information on the
differences between the US and Canada in terms of the
severity of the condition of patients admitted to hospital.
Even with these limitations, the NIRRU does give us
some advantages over a simple per diem, which does not
account for differences in resource-intensity during the
stay or for differences in severity of illness among
patients.
Assuming that Quebec’s hospitalizations represent
21% of those for Canada (based on the rate reported for
1989), the estimate by Krahn et al.14 for 1990, inflated at
3% per year (or $20 million), is very similar to the
estimate we have reached in this study. Their study used
a more macro approach, taking the proportion of total
Canadian hospital days accounted for by patients with
asthma and multiplying this by the aggregate cost data
for all Canadian public hospitals. Our research study
involved a micro approach, using the Quebec Ministry of
Health and Social Services reported cost and length of
stay by service and ward, in the majority of cases
adjusted specifically for the ward in which the patient
was treated. Nonetheless, the current study may have left
some asthma hospitalization costs unaccounted for, as
we reviewed only short-term hospital centres and only
hospitalizations for a period of 30 days or less. However,
those hospitalizations not included accounted for less
than 1% of the total asthma hospitalizations for 1994/95.
Conclusion
This study shows similar overall rates of
hospitalization in 1988/89 and 1994/95, but with an
important rise in the rates among very young children.
There were decreases in rates in very few age groups. A
small decrease in mean length of stay was seen.
1999
Many of the trends and variations in asthma hospital
use found in this study have been seen previously and
their causes discussed. Certainly, in analyses of
institutional database information it is difficult to
determine the causes of changing patterns of use.
However, even if the increases in rates of hospitalization
seen in very young children could be attributed in part to
a change in physician practice or an increase in the use
of asthma diagnosis, the rate has doubled since the
beginning of the decade and should signal an alarm that
justifies further investigation.
A one-year cost of $18–21 million may seem high for
asthma hospital costs, but it is most likely a considerable
underestimation of the true hospital cost because
emergency department visits and physician costs were
not included in the analysis.
Acknowledgements
Partial funding of this study was provided by Glaxo Canada
Inc. Dr Laurier holds the Hoechst Marion Roussel Chair on
Use of Medications: Policy and Outcomes at the University of
Montreal.
References
1.
2.
3.
4.
5.
Evans III R, Mullally DI, Wilson RW, Gergen PJ,
Rosenberg HM, Grauman JS, et al. National trends in the
morbidity and mortality of asthma in the US. Prevalence,
hospitalization and death from asthma over two decades:
1965S1984. Chest 1987;91(Suppl): 65SS73S.
Dodge RR, Burrows B. The prevalence and incidence of
asthma and asthma-like symptoms in a general population
sample. Am Rev Respir Dis1980;122:567S75.
Mak H, Johnston P, Abbey H, Talamo RC. Prevalence of
asthma and health service utilization of asthmatic children
in an inner city. J Allergy Clin Immunol 1982;70:367S72.
Émond A, Guyon L, Camirand F, Chenard L, Pineault R,
Robitaille Y. Et la santé, ça va? Tome 1, Rapport de
l’enquête Santé Québec 1987. Quebec: Publications du
Québec, 1988.
Mao Y, Semenciw R, Morrison H, MacWilliam L, Davies
J, Wigle D. Increased rates of illness and death from
asthma in Canada. Can Med Assoc J 1987;137:620S4.
6.
Boulet L, Milot J, Beaupré A. Mortalité associée à
l’asthme au Québec de 1975 à 1985. Union Médicale du
Canada 1989;118:150S7.
7. Wilkins K. Fact sheet: asthma (ICD-9 493). Chronic Dis
Can 1993;14(2):50.
8. Wilkins K, Y Mao. Trends in rates of admission to
hospital and death from asthma among children and
young adults in Canada during the 1980s. Can Med Assoc
J 1993;148:185S90.
9. Mao Y, Semenciw R, Morrison H, Wigle DT. Seasonality
in epidemics of asthma mortality and hospital admission
rates, Ontario, 1979S86. Can J Public Health
1990;81:226S8.
10. Osborne ML, Vollmer WM, Buist AS. Periodicity of
asthma, emphysema, and chronic bronchitis in a
87
northwest health maintenance organization. Chest
1996;110:1458S62.
11. Habbick B, Baker MJ, McNutt M, Cockcroft DW. Recent
trends in the use of inhaled B2-adrenergic agonists and
inhaled corticosteroids in Saskatchewan. Can Med Assoc
J 1995;153:1437S43.
12. Gerdtham U-G, Hertzman P, Johsson B, Boman G.
Impact of inhaled corticosteroids on acute asthma
hospitalization in Sweden. Med Care 1996;34:1188S98.
13. Grossman J. One airway, one disease. Chest 1997;111(2
Suppl):11SS16S.
14. Krahn M, Berka C, Langlois P, Detsky AS. Direct and
indirect costs of asthma in Canada, 1990. Can Med Assoc
J 1996;154:821S31.
15. Régie de l’Assurance maladie du Québec. Statistiques
annuelles. Quebec: Régie de l’Assurance maladie du
Québec, 1994.
16. Ministère de la Santé et des Services sociaux. Évaluation
de la performance économique globale des centres
hospitaliers de soins généraux et spécialisés, volet
“clientèle hospitalisée” — résultats 1994-1995.
Government of Quebec, 1996.
17. Ministère de la Santé et des Services sociaux, province de
Québec. Direction générale de l’administration et des
immobilisations [unpublished data]. Quebec, Canada.
18. Statistics Canada. Hospital statistics: preliminary annual
report, 1994S95. Health Rep 1996;8(1):48.
19. Johnston SL, Pattemore PK, Sanderson G, Smith S,
Campbell MJ, Josephs LK, et al. The relationship
between upper respiratory infections and hospital
admissions for asthma: a time-trend analysis. Am J Respir
Crit Care Med 1996;154(3 Pt 1):654S60.
20. Bassett IJ. Surveys of air-borne ragweed pollen in Canada
with particular reference to sites in Ontario. Can J Plant
Sci 1959;39:491S7.
APPENDIX
Example of the calculation of the NIRRU and adjustment
for length of stay (LOS)
NIRRUQ = NIRRUM + (NET LOSQ-M X NIRRUM ÷ LOSM X RatioFC)
where
NIRRUQ = Quebec cost index for the APR-DRG
NIRRUM = Maryland cost index for the APR-DRG
NET LOSQ-M = Difference between Quebec and Maryland of average length of stay for the APR-DRG
LOSM = Average length of stay for the APR-DRG in Maryland
Ratio FC = Ratio of fixed daily costs to average daily costs for the APR-DRG in Maryland
To calculate the NIRRU for an APR-DRG that, in Maryland, was given an index of 2.5 per case, had an average stay of 5 days
and a fixed-cost ratio of 60%,
NIRRUQ = 2.5 + ((7-5) X 2.5 ÷ 5 X 0.6) = 3.1
The last step is to normalize all the DRGs thus calculated to ensure that the total of the weighted cases equals the real total.
This calculation is applied to all except the atypical cases. Atypical cases are long-stay patients occupying short-stay beds,
patients who died, patients discharged without authorization, transfers, home-care patients, patients admitted and discharged
the same day, and patients whose stay exceeded a certain maximum (calculated to exclude about 3% of cases). NIRRUs for
atypical cases are calculated not on the basis of the average DRG but on their actual average stay in proportion to the average
Quebec stay for their DRG. An additional adjustment is also made to the NIRRU for those who have died or been transferred
according to the difference in the use of resources as a function of the date of death or transfer. O
88
Chronic Diseases in Canada
Vol 20, No 2
The Cost of Suicide Mortality in New Brunswick, 1996
Dale Clayton and Alberto Barceló
Abstract
Suicide is a major public health problem in Canada. Suicide deaths affect society by
consuming both human lives and economic resources. The present study estimates the
economic impact of suicide deaths that occurred in New Brunswick in 1996, using the human
capital approach. For the 94 suicide deaths reported, direct costs for health care services,
autopsies, funerals and police investigations were $535,158.32. Indirect costs, which estimate
the value of lost productivity due to premature death, had the largest economic value, of
$79,353,354.56. The mean total cost estimate per suicide death in 1996 was $849,877.80.
Although the most significant impact of a suicide death remains the loss of a human life, these
results indicate that the economic cost of this public health tragedy in New Brunswick is also
great. To our knowledge, this report provides the first complete cost-of-suicide analysis
performed in a Canadian province.
Key words: Canada; cost of illness; New Brunswick; suicide
Introduction
Suicide touches the lives of many Canadians. In
1996, suicide was the leading cause of death among
those aged 25S29 years in New Brunswick.1 Previous
studies have shown that suicide rates in New Brunswick
increased from 1987 to 1995,2 from 9.5 per 100,000
(95% confidence interval [CI] = 7.46S12.07) to 16.17
per 100,000 (95% CI = 13.48S19.36). Suicide deaths
accounted for 1.4% of all deaths occurring in New
Brunswick in 1987,3 and this figure rose to 2.0% of
all deaths in 1995,4 although the increase was not
statistically significant. This provincial ranking
compares with national statistics in which suicide
deaths remained at 1.9% of all deaths in both 19873
and 1995.4 An awareness of rising suicide rates has led
to questions about the economic impact of suicide in
New Brunswick.
Although many Canadian cost-of-illness studies
have been completed,5S11 no Canadian cost-of-suicide
studies have been reported that include both direct and
indirect costs. Earlier Canadian cost-of-suicide studies
have focused on only one category of costs (direct or
indirect), but not both. Hanvelt et al. estimated the
indirect cost per suicide death of men aged 25S64 in
Canada from 1987 to 1991 at $516,800 (1990 US$).11
Using the conversion rate to 1990 Canadian dollars cited
by Hanvelt et al. of $1.167 (CAN) = $1 (US) and
adjusting to 1996 Canadian dollars using the consumer
price index,12 the estimate by Hanvelt et al. becomes
$684,524.86 per suicide death. Hanvelt et al. further
reported that for men of this age group, suicide deaths
resulted in the third greatest loss in productivity, behind
HIV/AIDS deaths and motor vehicle fatalities.11
Similarly, Miller estimated suicide deaths by gunshot
in Canada in 1991 to cost $1,036,494 per suicide death
(1993 CAN$);6 this does not include Miller’s estimate
of lost quality of life. Adjusting to 1996 dollars, this
estimate translates to $1,078,238.85 per suicide death.
Several US studies have examined the economic
burden of suicide deaths in association with specific
mental disorders13S15 or within specific age groups.16
Although these studies have been useful in quantifying
the cost of suicide within these groups, the appropriateness of extrapolating from these studies to a broader
Canadian population remains questionable. Issues that
arise when making these US–Canadian comparisons
include differences in health care costs, given the
different health care systems in the two countries, and
the socio-economic representativeness of suicides
occurring within specific age or mental illness categories
with respect to all suicides occurring in Canada. Table 1
summarizes the results of previous cost-of-illness studies
as they relate to suicide deaths.
Author References
Dale Clayton, New Brunswick Department of Health and Community Services, PO Box 5100, Fredericton, New Brunswick E3B 5G8; E-mail:
[email protected]
Alberto Barceló, Pan-American Health Organization, Regional Office of the World Health Organization, Washington, DC, USA
1999
89
TABLE 1
Previous cost-of-illness studies including data on suicide deaths
Author
Country
Year
Cost per suicide death
Comments
Stoudemire et al.14
US
1980
$260,691
(1980 US$)
Includes only discounted lifetime earnings
Hanvelt et al.11
Canada
1987–1991
$516,800
(1990 US$)
Includes only production losses for suicide deaths of males
aged 25–64
Miller6
Canada
1991
$1,036,494
(1993 CAN$)
Does not include Miller’s estimate of “lost quality of life;”
based only on deaths by gunshot
Wyatt and Henter15
US
1991
$591,475
(1991 US$)
Includes indirect costs only
Palmer et al.13
US
1994
$397,066
(1994 US$)
Includes indirect and direct costs; incidence was estimated
by applying 1991 rates to 1994 population
This study used an incidence-based human capital
approach to estimate the total economic burden of
suicide deaths in New Brunswick in 1996.
Methods
Identifying Completed Suicide Cases
Suicide deaths that occurred in New Brunswick in
1996 were identified from the New Brunswick Vital
Statistics database. Name, address, Medicare number,
date of birth and date of death were collected for each
case, and each case was cross-referenced with the suicide
database of the New Brunswick Coroner’s Office to
ensure that all reported suicide deaths in the province
had been identified.
Cost Estimates
Costs were conceptually categorized into two groups:
actual dollar expenditures related to the suicide death
(direct costs) and estimates of the value of future
productivity losses (indirect costs). It should be noted
that this classification into direct and indirect categories
is arbitrary and does not affect the resulting economic
value estimates.
Direct Costs
Ambulance services
The use of ambulance services in association with
each suicide case was determined by cross-referencing
Medicare numbers with the New Brunswick Ambulance
Services database. The average cost per ambulance
service in 1996 was then applied to the identified cases.
This cost was estimated by dividing the total cost of
operating ambulance services in the province during the
1996/97 fiscal year by the number of service calls
answered during that period.
Hospital services
The Hospital Financial Utilization Management
System (HFUMS) of the Department of Health and
Community Services was cross-referenced with
90
Chronic Diseases in Canada
Medicare numbers of completed suicide cases. Hospital
services that were shown to be associated with a suicide
attempt and subsequent death were included. For each
case identified, the records department of the associated
hospital was contacted and asked for the Resource
Intensity Weight (RIW) assigned to each case by the
Canadian Institute for Health Information.17 Using the
most recent dollar value equivalent (1994/95) calculated
for HFUMS adjusted to 1996 values according to the
consumer price index18 (1 RIW = $2,829.62), actual cost
estimates for each case were made by multiplying the
RIW by the equivalent dollar value.
Physician services
Costs for physician services were identified by
cross-referencing Medicare numbers of completed
suicides cases with the New Brunswick Medicare
payment database. Physician services provided one day
before a successful suicide attempt and up to and
including the day of death were included in our cost
estimate.
Autopsy services
Data relating to costs of autopsy services were
obtained directly from the New Brunswick Coroner’s
Office. These data reflected costs associated with
performing a post-mortem examination but did not
include such services as transporting the body to a
regional morgue facility.
Funeral/cremation services
A provincial average cost for funeral and cremation
services provided in 1996 was calculated from
information obtained from all funeral/cremation service
providers in the province. All service providers were
asked to report the number of funeral/cremation services
provided by their establishment in 1996 as well as the
average cost for those services. A provincial average was
then calculated by weighting the information supplied by
each provider to the total number of services that were
reported.
Vol 20, No 2
Police investigations
Costs incurred for the police investigation of suicide
deaths in New Brunswick in 1996 were obtained directly
from all police forces in the province. Each of the 16
regional police forces as well as the New Brunswick
Division of the Royal Canadian Mounted Police were
asked to supply the number of suicide cases investigated
in their jurisdiction during the 1996 calendar year. Each
police force provided estimates of the total cost incurred
for their suicide investigations based on personnel costs.
Indirect Costs
Potential years of life lost (PYLL)
PYLL calculations provide a quantifiable estimate of
the number of years of potential or productive life lost as
a result of premature death. PYLL before age 75 were
calculated and expressed as a ratio to the mean
population under age 75 in 1996, as described by the
Canadian Institute for Health Information.19
Discounted future earnings (DFE)
Lost future earnings as a result of suicide death were
calculated using the model developed by Vodden et al.
for the Ontario Ministry of Transportation.10 The model
estimates total future earnings lost as a result of
premature death for both labour force work and unpaid
labour (see Appendix). Future earnings were discounted
by an appropriate rate to convert them to present day
values. A rate of 4% was used for calculations based on
recommendations made by Miller et al.20 and the
experiences of other cost-of-suicide reports.13 All
calculations of discounted future earnings were carried
out using the S-Plus version 4.5 software package.21
Survival probabilities, Py,s(n), were collected from the
New Brunswick Statistics Agency based on published
reports from Statistics Canada.22 Mean annual
employment earnings by age and sex in New Brunswick,
Ys(n), were obtained from the 1996 Statistics Canada
Labour Force Survey.23
Information relating to the value of homemaking
services was taken from the Statistics Canada publication
Households’ Unpaid Work: Measurement and
Valuation.24 Data drawn from this publication were
based on the results of the 1992 national General Social
Survey (GSS) using a generalist approach for estimating
replacement costs (see reference for a detailed
description of methodology). Replacement cost data by
demographic group were converted to annual dollars per
person using census population estimates.24 The value of
homemaking services by age and sex, Yhs(n), was then
entered into the DFE model similarly to mean annual
employment earnings, Ys(n) (see Appendix).
Labour productivity data are reported annually by the
Centre for the Study of Living Standards.25 A rate of
1999
increase in labour productivity of 0.04
$GDP/hour/employed worker was estimated by
calculating the mean of the three-year moving averages
of the annual change in labour productivity from 1984 to
1995.
Results
Ninety-three suicide deaths were identified in both the
Vital Statistics and Coroner’s Office databases. One
additional case was found in the Coroner’s Office
database that did not appear in the Vital Statistics data.
Information from both sources was combined, and a total
of 94 deaths was used in all calculations. Table 2
summarizes the total cost estimates associated with
suicide deaths in New Brunswick in 1996. The total cost
of suicide deaths was calculated to be $79,888,513.17.
TABLE 2
Costs associated with suicide deaths, by cost
category, New Brunswick, 1996
Cost category
n
Cost
Calculation
method
DIRECT COSTS
Ambulance services
51
$38,130.00
Average
Hospital services
6
$35,130.87
Actual
Physician services
Actual
19
$4,393.67
Autopsy services
74
$26,000.00
Average
Funeral services
94
$380,358.78
Average
Police investigations
69
$51,145.00
Average
Discounted future earnings
94
$79,353,354.56
Estimate
TOTAL
94
$79,888,513.17
INDIRECT COSTS
Direct Costs
Ambulance services
Sixty-two ambulance services were provided to 51
suicide cases in 1996. Multiple services were provided
to some cases for transfer between hospitals. The total
cost for the year was calculated by applying the average
cost per ambulance service call of $615 to the 62 cases
($615 x 62 cases = $38,130.00).
Hospital services
A search of the Hospital Services database identified
six completed suicide cases in 1996 that had received
hospital services as a result of the suicide. For each of
these cases, RIW values ranged from 0.6430 to 4.0404.
Applying dollar value equivalents to these RIW values
resulted in a range of costs from $1,819.45 to
$11,432.80. The total cost for hospital services for
these six cases was $35,130.87.
91
Physician services
Physician services costs were included where
payment for such services was made by New Brunswick
Medicare. A search of the Medicare database found
physician services paid for 19 of the 94 suicide cases.
Six of these nineteen cases included physician services
provided during a hospital stay. Payment for physician
services by case ranged from $15.30 to $1,098.33, with a
total cost of $4,393.67 for all cases in 1996.
Autopsy services
New Brunswick Coroner Services recorded 74
autopsies for suicide-related deaths that occurred in
1996. A total cost of $26,000.00 was reported for the
provision of these services, at an average cost of $351.35
per suicide death.
Funeral/cremation services
the 94 suicide deaths in 1996 were calculated to be
$79,353,354.56, providing the largest contributing factor
(99%) of the overall cost of suicide deaths. A sensitivity
analysis of the 4% discount rate was performed to assess
the variability of discounted future earnings at 2%, 4%,
6%, 8% and 10% discount rates. The results of this
analysis are presented in Figure 1.
Method of Suicide
Since average costs were used in the estimation of
ambulance, autopsy and funeral services, and police
investigation costs were not available by suicide method,
accurate comparison of total costs by suicide method is
not possible. However, including only cost estimates for
hospital and physician services provides a relative
ranking of suicide method by health care costs (Table 4).
Using this comparison, suicide by firearm was the most
costly method ($13,920.32), followed by jump/fall
($12,531.13) and strangulation ($10,541.85).
Responses from individual funeral/cremation service
providers recovered cost information on 3,507 services
out of a total of 5,949 services provided in the province
in 1996.26 Based on this 59% of all services provided, an
average cost of $4,046.37 was calculated. Applying this
average cost to each of the 94 cases occurring in 1996
resulted in a total cost of $380,358.78.
Age
group
Police investigations
0–9
All the 17 police forces in New Brunswick
participated in the current study. A total of 69
investigations were reported across the province with
costs ranging from $180 to $2,500 per case. A total of
$51,145.00 was spent on police investigations of suicide
deaths, with an average cost of $741.23 per case.
Indirect Costs
Potential years of life lost (PYLL)
Using an upper age limit of 75 years,
potential
years of life lost were calculated for all
suicide cases occurring in 1996. Total
PYLL by age and sex are shown in
Table 3 as well as PYLL rates per 1,000
persons. The largest number of PYLL
occurred among those aged 35 to 49.
Total PYLL for men of this age group
was 1,227 years and for women, 170
years. The second largest loss was seen
among those aged 20 to 34 (852 and 148
years for men and women respectively).
TABLE 3
Potential years of life lost (PYLL) due to suicide
deaths, by age and sex, New Brunswick, 1996
Males
n
Females
PYLL
n
BOTH SEXES
PYLL
n
PYLL
0
0
0
0
10–19
5
299
2
119
7
418
20–34
18
852
3
148
21
1000
35–49
37
1227
6
170
43
1397
50–64
12
231
2
41
14
272
8
24
1
8
9
32
80
2633
14
486
94
3119
65+
TOTAL
Per 1000
7.21
0
1.34
0
4.29
FIGURE 1
Total discounted future earnings (DFE) by discount rate
Discounted future earnings
Total discounted future earnings were
calculated for each case in 1996 using a
discount rate of 4% (see Appendix).
Estimated discounted future earnings for
92
Chronic Diseases in Canada
Vol 20, No 2
The greatest cost of suicide deaths occurred in those
age groups with the greatest number of suicide deaths.
This was found to be the 35S49 age group for both men
($31,805,592.92; n = 37) and women ($4,418,836.70;
n = 6) (Table 5). Of the 94 suicide deaths occurring in
1996, 85.1% were male, accounting for 84.4% of the
total provincial cost of suicide.
Discussion
The results of the present study indicate that suicide
deaths have a significant impact on the New Brunswick
economy. Total average cost per suicide death in 1996
was estimated to be $849,877.80. The per capita cost of
suicide deaths for the population of New Brunswick was
$104.84. This does not include the emotional and
psychological burden experienced by the friends and
family members of suicide victims. Nor does it
encompass the value of that part of a person’s life that
cannot be estimated simply by a loss in productivity
(e.g. the value of being someone’s friend, the value of
TABLE 4
Hospital and physician service costs
associated with suicide deaths, by method,
New Brunswick, 1996
Suicide
methoda
(ICD-9
code)
Hospital
services
(n)
Firearm
(955.0–955.4)
$13,314.21
(2)
50–64
12
Physician
services
(n)
TOTAL COST
being aware of one’s existence, the value of emotional
experiences, etc.).
This study made use of the human capital approach to
estimate the value of lost productivity due to premature
death. This approach estimates loss by approximating
current market values for lost productivity in the
future.4,9,27 This estimate is made for both paid (e.g.
employment) and unpaid labour (e.g. homemaking). Lost
productivity can be quite easily summed using widely
accepted numerical values; employment earnings by age
and sex are easily reproducible and measurable numbers.
Using a discounted future earnings model,10 the age- and
sex-specific values of both paid and unpaid labour can be
summed over the years of life lost to premature death,
and adjusted by labour productivity and discount rates.
Another method of approximating the value of life
lost is the “willingness-to-pay” approach.4,9,27 Unlike the
human capital approach, which estimates the market
value of human productivity, willingness-to-pay reflects
the societal value of life by estimating the amount of
money people would be willing to pay to avoid a suicide
death. The willingness-to-pay approach is believed to
assign a greater economic value to lost life than the
human capital approach, as it encompasses the
psychological and physical burden of pain, suffering and
lost quality of life. The decision to use the human capital
approach in this study was made on the basis of the
availability of relevant data sources and the
reproducibility of the study’s results.
This study provides a reference point from which
subsequent cost-of-suicide studies may be compared.
These results are similar to those reported by Hanvelt et
Jump/fall
$11,432.80
$1,098.33
$12,531.13
al.11 and Miller,6 given their focus on either partial cost
(957)
(1)
(1)
or specific method. Indirect costs per suicide death of
males aged 25S64 estimated by Hanvelt et al. were
Strangulation
$8,564.41
$1,977.44
$10,541.85
(953.0–953.9)
$684,524.86 (adjusted to 1996$). As expected, this was
(2)
(8)
less than our estimate of both indirect and direct costs for
Overdose
$1,819.45
$711.79
$2,531.24
both sexes and all age groups, of $849,877.80. Miller
(950.0–950.9)
(1)
(4)
estimated total costs per suicide death by gunshot to be
TOTAL
$35,130.87
$4,393.67
$39,524.54
$1,078,238.85 (adjusted to 1996$). This does not include
Miller’s estimate of “lost quality of
life”, which was not included in the
TABLE 5
present study. Miller’s estimate was
greater than that for the gunshot
Total cost of suicide deaths, by age and sex, New Brunswick, 1996
suicides included in our study
($816,849.51 per death). This
Age group
Males
Females
BOTH SEXES
difference is likely due to estimates
n
Cost
n
Cost
n
Cost
of discounted future earnings in the
0–9
0
—
0
—
0
—
two studies. The age and sex of the
10–19
cases sampled heavily influence
5
$7,131,343.41
2
$2,877,343.60
7
$10,008,687.0
1
discounted future earnings estimates.
20–34
18
$22,528,517.8
3
$3,766,926.95
21
$26,295,444.7
As may have occurred in Miller’s
2
7
study, a sample of younger cases
35–49
37
$31,805,592.9
6
$4,418,836.70
43
$36,224,429.6
with a higher proportion of males
1
1
would result in a higher estimate.
1999
$4,814,204.44
$606.11
(6)
$13,920.32
2
$1,093,031.02
14
$5,907,235.46
93
This report provides an overall cost estimate of nearly
$80 million for suicide deaths occurring in New
Brunswick in 1996, which translates to an average cost
of $849,877.80 per suicide death. With greater
refinement of data sources, future studies should include
the cost of mental health services for friends and family
members of suicide victims and an estimation of lost
productivity values for those grieving a suicide death.
It would also be possible to broaden the scope of
future cost-of-suicide studies to include all forms of
suicidal behaviour. This would include not only costs
following a suicide death, but costs incurred for each
suicidal person before their successful attempt. A further
step would be to then include costs associated with
suicide attempts that never result in suicide death. Such a
study would provide a complete profile of the economic
impact of suicidal behaviour in society. Nevertheless, the
present study provides the groundwork for further
research.
Summary
In summary, this report describes the costs associated
with suicide deaths occurring in New Brunswick in
1996. The model used to assess these costs involved an
incidence-based human capital approach involving both
direct and indirect costs. A total cost of $79,888,513.17
was reported for the 94 suicide deaths in 1996, resulting
in a mean cost of $849,877.80 per suicide death. An
itemized breakdown of costs was presented by age, sex
and method of suicide. Comparisons with other
Canadian cost-of-fatal illness studies are discussed, and
suggestions for further research are given.
Acknowledgements
The authors wish to thank Varsha Chhatre, biostatistician with
the New Brunswick Provincial Epidemiology Service, for her
technical assistance in the calculation of discounted future
earnings.
References
1.
New Brunswick Department of Health and Community
Services. Annual statistical report, 1996. Fredericton:
Vital Statistics, 1996.
2.
New Brunswick Department of Health and Community
Services. Suicide in New Brunswick, 1987S1995.
Fredericton: Provincial Epidemiology Service, 1997 Nov.
Statistics Canada. Mortality: summary list of causes.
Ottawa, 1989; [formerly] Cat 84-206.
Statistics Canada. Causes of death, 1995. Ottawa, 1997;
Cat 84-208-XPB.
Moore R, Mao Y, Zhang J, Clarke K. Economic burden of
illness in Canada, 1993. Ottawa: Canadian Public Health
Association, 1997.
3.
4.
5.
6.
Miller T. Costs associated with gunshot wounds in
Canada in 1991. Can Med Assoc J 1995;153:1261S8.
7.
Kaiserman MJ. The cost of smoking in Canada, 1991.
Chronic Dis Can 1997;18(1):13S9.
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Chronic Diseases in Canada
8.
Chan B, Coyte P, Heick C. Economic impact of
cardiovascular disease in Canada. Can J Cardiol
1996;12(10):1000S6.
9. The Canadian Burden of Illness Study Group. Burden of
illness of multiple sclerosis. Part I: cost of illness. Can J
Neurol Sci 1998;25(1):23S30.
10. Vodden K, Meng R, Smith D, et al. The social cost of
motor vehicle crashes, final report to Ontario Ministry of
Transportation. Ottawa: Abt Associates of Canada, 1993.
11. Hanvelt R, Ruedy N, Hogg R, Strathdee S, Montaner J,
O’Shaughnessy M, Schechter M. Indirect costs of
HIV/AIDS mortality in Canada. AIDS 1994;8:F7S11.
12. Statistics Canada. Canadian economic observer,
historical statistical supplement 1997/98. Ottawa,
1998; Cat 11-210-XPB.
13. Palmer C, Revicki D, Halpern M, Hatziandreu E. The cost
of suicide and suicide attempts in the United States. Clin
Neuropharmacol 1995;18(3):S25S33.
14. Stoudemire A, Frank R, Hedemark N, Kamlet M, Blazer
D. The economic burden of depression. Gen Hosp
Psychiatry 1986;8:387S94.
15. Wyatt R, Henter I. An economic evaluation of
manic-depressive illness, 1991. Soc Psychiatry Psychiatr
Epidemiol 1995;30:213S9.
16. Weinstein M, Saturno P. Economic impact of youth
suicides and suicide attempts. In: Report of the
Secretary’s Task Force on Youth Suicide. US
Government, 1986:4-82S4-93.
17. Johnson L, Richards J, Pink G, Campbell L. Case mix
tools for decision making in health care. Ottawa:
Canadian Institute for Health Information, 1998.
18. Statistics Canada. Consumer price index, 1996
classification, annual average indexes, Canada and
provinces. CANSIM, matrix 9961. Ottawa.
19. Working Group on Community Health Information
Systems and Chevalier S, Choinière R, Ferland M,
Pageau M, Sauvageau Y. Community health indicators.
Definitions and interpretations. Ottawa: Canadian
Institute for Health Information, 1995.
20. Miller T, Whiting B, Kragh B, Zegeer C. Sensitivity of
resource allocation models to discount rate and
unreported accidents. Transportation Research Record
1987;1124:58S65.
21. S-Plus version 4.5 [computer software]. Cambridge
(MA): MathSoft, Inc. 1995.
22. Statistics Canada. Population projections for Canada,
provinces and territories, 1993-2016. Ottawa, 1995; Cat
91-520.
23. Statistics Canada. Labour force estimates by detailed age,
sex, New Brunswick, 1996 (Table 01AN). Labour Force
Historical Review (71F0004XCB). Ottawa, 1997.
24. Statistics Canada. Households’ unpaid work:
measurement and valuation. Ottawa, 1995; Cat
13-603-MPE no 3.
Vol 20, No 2
25. Capital, labour and total factor productivity tables by
province. Ottawa: Centre for the Study of Living
Standards, 1997.
26. New Brunswick Department of Health and Community
Services, Vital Statistics [unpublished 1996 data].
27. Rice DP, MacKenzie EJ, et al. Cost of injury in the United
States: a report to Congress. San Francisco (CA):
Institute for Health & Aging, University of California,
and Injury Prevention Center, The Johns Hopkins
University; 1989.
APPENDIX
Discounted future earnings model
99
DFE = 3 Py,s(n) [Ys(n) Es(n) + Yhs(n) Ehs(n)] x (1+g)n-y
n=y
(1+r))n-y
where
DFE = discounted future earnings lost due to suicide death
Py,s(n) =
probability that a person of age y and sex s will survive to age n
y=
age at which the individual successfully completed suicide
s=
sex of the individual
n=
age of the individual
Ys(n) = mean annual earnings of an employed person of sex s and age n including the value of fringe
benefits
Es(n) = proportion of the population of sex s and age n that are employed in the labour market
Yhs(n) = mean annual imputed value of homemaking services of a person of sex s and age n
Ehs(n) = proportion of the population of sex s and age n that are keeping house
g=
rate of increase of labour productivity
r=
real discount rate
Source: Adapted, with permission, from Reference 10 O
1999
95
Workshop Report
Canadian National Workshop on Measurement of
Sun-related Behaviours
Chris Lovato, Jean Shoveller, Christina Mills and an Expert Panel
Introduction
Skin cancer has been described as an emerging
epidemic in North America.1–3 The National Cancer
Institute of Canada estimates that 66,000 new cases of
skin cancer will be diagnosed in 1999 in Canada.4
Epidemiologic evidence suggests that cumulative
exposure to sunlight is necessary for the development of
squamous cell carcinoma, whereas intense sun exposure
and sunburns received during childhood are more
important in the development of melanoma and basal
cell carcinoma of the skin.5–9
Sun exposure and protection are at least partly under
the control of individuals. Therefore, it is important to
pursue behavioural interventions as an essential
component of a comprehensive set of skin cancer
prevention strategies. In turn, monitoring sun-related
behaviours is important in the development and
evaluation of interventions as well as in surveillance of
behavioural change across populations. One of the
greatest challenges facing behavioural researchers in
skin cancer prevention is the lack of comparability of
measures across studies. The identification of a standard
set of items would enhance the quality of research and
program evaluation in Canada and contribute to similar
efforts in other parts of the world.
The idea for the workshop described in this paper
emerged from recommendations identified at the 1997
Workshop on Research, Policy and Program Planning on
Sun Protective Behaviours.10 At the 1997 workshop,
participants developed a set of recommendations for
research initiatives related to sun protection efforts in
Canada. One of the primary recommendations was to
develop a standard set of definitions and core items to
assess sun-related behaviours. In response, Chris Lovato,
Jean Shoveller and Christina Mills agreed to co-chair the
1998 Canadian National Workshop on Measurement of
Sun-Related Behaviours. The purpose of this workshop
was to develop consensus on a standard set of measures
for program evaluation and for monitoring of sun
exposure and protective behaviours in Canada.
Before the workshop took place, a systematic review
of the published literature was conducted to collect
reports and instruments previously used to measure
sun-related behaviours. The search of electronic
databases (e.g. MEDLINE, CancerLit) included
peer-reviewed and non-peer-reviewed publications and
was limited to English-language articles published
between 1990 and 1998. In addition, the reference lists
of each of the publications retrieved were scanned.
Members of the workshop planning committee also
recommended published reports and unpublished
instruments for inclusion. Finally, scientists working in
the field were asked for copies of instruments that had
not been published previously. As a result, a total of 112
publications and instruments were identified.
Five background papers were commissioned, which
synthesized and critically assessed the literature and
available instruments in the areas of sun exposure,
sunburn, protective behaviours, artificial tanning and
phenotype. Each background paper provided
recommendations regarding standardized operational
definitions and core items that could be used in routine
behavioural surveillance efforts and program evaluation.
Participants were provided with the background papers
before the workshop.
Author References
Chris Lovato and Jean Shoveller, Centre for Community Child Health Research, Department of Health Care and Epidemiology, University of British
Columbia
Christina Mills, Cancer Bureau, Laboratory Centre for Disease Control, Health Canada, Ottawa, Ontario
Expert Panel: List of workshop participants at end of article
Correspondence: Dr Chris Lovato, Associate Professor, Centre for Community Child Health Research, L408 – 4480 Oak Street, Vancouver, British
Columbia V6H 3V4; Fax: (604) 875-3569; E-mail: [email protected]
96
Chronic Diseases in Canada
Vol 20, No 2
Summary of Workshop Proceedings
The workshop was held on October 29–30, 1998, at
the University of British Columbia. Invited participants
(17 in total) included skin cancer researchers and
representatives from Health Canada, Environment
Canada, the Canadian Dermatology Association and the
Canadian Association of Optometrists. A representative
from the US Centers for Disease Control and Prevention
was also invited to attend. Participants engaged in a
series of small group discussions about the background
papers and identified specific recommendations for
operational definitions and core items. These
recommendations focused on measurement issues related
to sunburn, phenotype, sun exposure and protective
behaviours. Participants also identified priorities for
further research on measurement of sun-related
behaviours. Two themes emerged over the two-day
meeting: the unique characteristics of sun-related
behaviours and the challenges associated with data
collection and measurement.
Participants identified a number of factors that
differentiate sun-related behaviours from other health
behaviours (e.g. those related to nutrition, tobacco).
• In Canada, sun exposure is by and large a seasonal
behaviour that varies geographically as a result of the
large land mass and differences in weather patterns.
Thus, a national study conducted over a broad
geographic area that examines a specific exposure
period requires measures sensitive to such external
influences.
• Sun-related behaviours require individuals to interpret
and respond to their risk according to a complex set of
environmental and physiological cues.
• Comprehensive sun protection requires individuals to
undertake a set of behaviours. Therefore, research
must account for a set of behavioural outcomes rather
than a single indicator.
• A number of non-behavioural factors influence
individual risk, including phenotype, occupation and
age.
• Some sun protection messages (e.g. avoid sun
between 11 am and 4 pm) may conflict with health
messages that promote participation in outdoor
physical activities.
• There is a lack of definitive evidence regarding the
effectiveness of sunscreen, although it is one of the
primary methods of sun protection used by many
people.
1999
In discussing the background papers and the unique
nature of sun-related behaviours, workshop participants
also identified a number of points related to data
collection and measurement.
• Most previous research has relied on self-reported
behaviours by youth and adults and proxy reports for
children. The limitations of self-report and proxy data
should not be overlooked. Some recent studies have
tried to incorporate objective data collection tools
(e.g. daylight exposure monitors worn as
wristwatches), which may serve as useful approaches
to developing validated self-report measures.
• Operational definitions of sun-related variables have
varied greatly across published studies. This was
identified as a serious barrier to advancing this area of
research.
• There is also considerable variation related to recall
periods. For example, some studies assess general
patterns of sun exposure and protection (e.g. over the
entire summer), whereas others assess behaviours in
specific, brief time periods (e.g. the previous
weekend).
• More work is needed to develop scales and indices for
assessment of sun-related behaviours. Most previous
studies have reported on individual protective
behaviours and have not considered the potential
cumulative effect of multiple protective behaviours.
The workshop resulted in a set of recommendations
that will be useful to those conducting research and
program evaluations in this area. Overall, workshop
participants identified measurement research as a high
priority.
Recommendations
Core Items
Six core items were developed for inclusion in
omnibus-style behaviour surveillance surveys and
smaller scale evaluation efforts. Table 1 lists the
recommended core items in order of priority—sunburn,
phenotype, sun exposure and sun protection—and
summarizes the rationale for each item. These core items
are recommended for inclusion in population-based
surveys, including the Canadian National Population
Health Survey. The items were developed for use in
personal interviews, telephone interviews or
self-administered survey formats. They are suitable for
wider dissemination within Canada and internationally to
other researchers in skin cancer prevention.
97
TABLE 1
Recommended core items for measuring sun-related behaviours
Item
Comment
SQ1: A sunburn is any reddening or discomfort of your skin that lasts
longer than 12 hours after exposure to the sun or other UV
[ultraviolet] sources, such as tanning beds or sunlamps. In the
past
year, has any part of your body been sunburned?
Yes/No
Sunburn
These three items have been chosen as the most
important items to be asked in an omnibus survey
because they measure sunburn and also provide
indirect measures of sun exposure and protective
behaviours. If there is limited space, this series of
items should be used.
Universe: All respondents
SQ2: Did any of your sunburns involve blistering?
Yes/No
Universe: Respondents who had a sunburn in previous year
“Past year” was identified as the time frame for
reporting because sunburns are not of high
frequency or routine events. Use of this term also
allows for ease of administering the item during any
time of the year.
SQ3: Did any of your sunburns involve pain or discomfort that lasted
for
more than one day?
Yes/No
Universe: Respondents who had a sunburn in previous year
SQ4: Would you say the untanned skin colour of your inner
upper arm is ...
Light (white, fair, ruddy)
Medium (olive, light brown, medium brown)
Dark (dark brown, black)
Phenotype
Phenotype is one of the primary risk factors
associated with skin cancer. This item allows for
more precise interpretation of the data collected
using the above sunburn items.
Universe: All respondents
SQ5: During this past June through August, on a typical weekend or
day off from work, approximately how much time did you spend in the
sun between 11 am and 4 pm?
<30 minutes per day
30 minutes to 1 hour per day
1–2 hours per day
2+ hours per day
Sun exposure
Sun exposure is a risk factor associated with skin
cancer. In Canada, peak UVB exposure from the sun
occurs during June through August between the
hours of 11 am and 4 pm.
Universe: All respondents
SQ6: Think of the most recent weekend or day off from work when
you spent 30 or more minutes in the sun. Did you ...
Seek shade: Yes/No
Wear a hat that shaded your face, ears and neck: Yes/No
Wear a shirt with long sleeves: Yes/No
Wear long pants or a long skirt: Yes/No
Use sunscreen with SPF 15+ on all exposed skin: Yes/No
Universe: Respondents who spent 30+ minutes per day in the sun
98
Chronic Diseases in Canada
Sun protection
To maximize the accuracy of recall, this item should
be used only if the survey is administered between
the beginning of June and the end of September.
The Canadian Dermatology Association has
identified these five actions to protect against skin
damage during exposure to the sun for 30+
minutes. All five actions are endorsed in the
consensus statement generated at the National
Workshop on Public Education Messages for
Reducing Health Risks from Ultraviolet Radiation.
Vol 20, No 2
Further Research
The recommended core items should be pilot tested to
assess their validity and reliability. In general, validation
studies of self-report items should be considered a
priority—for example, studies to validate self-reported
phenotype against objective measures of skin colour and
self-reported sun exposure against the results of daylight
exposure monitors. Scales and indices that provide a
composite score representing exposure and protective
behaviours should be developed and validated.
International Collaboration
To advance the quality of measurement of sun-related
behaviours and expand knowledge in this area of
ultraviolet radiation research, an international
collaboration to develop consensus regarding operational
definitions and core items should be undertaken. On the
basis of the success of the 1998 Canadian National
Workshop on Measurement of Sun-Related Behaviours,
it is recommended that this approach be considered for
an international meeting.
Discussion
As skin cancer continues to be a public health
problem in North America, surveillance will be required
to monitor the prevalence of sun-related behaviours at
the population level. As communities continue to
demand interventions to prevent skin cancer, there will
be an increasing need to evaluate the efficacy of
programs. The recommended core items here are based
on the accumulated evidence currently available. There
are limitations to these items, including the accuracy of
self-reported data. Further, the items developed during
the workshop focused on adolescent and adult
populations. Since reduction of exposure to UVB during
childhood is critical to reducing the risk of skin cancer,
measures need to be developed for use with this age
group (e.g. parental proxy reports). New strategies also
need to be developed to collect data from children
themselves. The core items focus on behaviour and do
not attempt to assess attitudes or barriers; local program
evaluations may need to include supplemental items to
address these areas. Finally, the survey items have not
been tested and will require further assessment.
The process used in this workshop was influenced by
previous efforts to establish standardized measures for
use in tobacco control research.11 The background papers
developed and disseminated to participants before the
workshop facilitated a common understanding of the
conceptual and methodologic issues. The workshop
process could also be applied to other areas, such as
measurement of nutrition and physical activity.
that although there is an established body of research in
sun protection, it is extremely difficult to compare results
across studies because of the wide variations in the way
behaviours are measured. Common barriers to evaluating
interventions are the availability of measures and
standards of acceptability. The establishment and use of
core items help to address both issues.
Progress is being made regarding implementation of
the workshop recommendations. The core items
presented in Table 1 are currently being considered for
inclusion in the Canadian National Population Health
Survey. The items have also been presented at the 1999
biennial meeting of the Canadian Society of
Epidemiology and Biostatistics. A number of workshop
participants are conducting research in the following
areas: pilot testing the core items, validating
self-reported exposure and skin colour, and constructing
an index for sun exposure and protective behaviours. In
addition, researchers are developing items for use in
measuring parents’ reports of their children’s sun-related
behaviours.
These recommended core items will be useful to
researchers and program evaluators addressing
sun-related behaviours. Researchers should continue to
conduct measurement studies in order to improve the
quality of surveillance and evaluation tools. Efforts are
currently under way to promote international consensus
regarding measurement.
Acknowledgements
The workshop was supported by funding from Health Canada
and the Terry Fox Workshop Program, which is administered
by the National Cancer Institute of Canada.
References
1.
2.
3.
4.
5.
6.
To advance the development of knowledge related to
health behaviour change, more comparability across
research studies is required in Canada as well as
internationally. In organizing this workshop, we noted
1999
Gallagher RP, Hill GB, Bajdik CD, et al. Sunlight
exposure, pigmentary factors, and risk of nonmelanocytic
skin cancer. I. Basal cell carcinoma. Arch Dermatol
1995;131:157–63.
Gallagher RP, Hill GB, Bajdik CD, et al. Sunlight
exposure, pigmentary factors, and risk of nonmelanocytic
skin cancer. II. Squamous cell carcinoma. Arch Dermatol
1995;131:164–9.
Rivers JK. Melanoma. Lancet 1996;347:803–7.
National Cancer Institute of Canada. Canadian cancer
statistics 1999. Toronto: NCIC, 1999.
Gallagher RP, MacLean DI, Yang P, Coldman AJ, Silver
HK, Spineli JJ, Beagrie M. Suntan, sunburn, and
pigmentation factors and the frequency of acquired
melanocytic nevi in children. Similarities to melanoma:
the Vancouver mole study. Arch Dermatolol
1990;126:770–6.
Rosso S, Zanetti R, Martinez C, et al. The multicentre
south European study ‘Helios’: different sun exposure
patterns in the etiology of basal cell and squamous cell
carcinoma of the skin. Br J Cancer 1996;73:1447–54.
99
7.
Weinstock MA, Colditz GA, Willett WC. Nonfamilial
cutaneous melanoma incidence in women associated with
sun exposure before 20 years of age. Pediatrics
1989;84:199–204.
8. Gibbons L, Anderson L, eds. Proceedings of the
Symposium on Ultraviolet Radiation-related Diseases.
Chronic Dis Can 1992;13(5 Suppl): S1–42.
9. Mills CJ, Trouton KJ, Gibbons L. Symposium report:
Second Symposium on Ultraviolet Radiation-related
Diseases. Chronic Dis Can 1997;18(1):27–38.
10. Ashbury FD, Rootman I. Workshop report: research,
policy and program planning on sun protective
behaviours. Cancer Prev Control 1998;2(3):129–32.
11. Mills C, Stephens T, Wilkins K. Workshop report:
summary report of the Workshop on Data for Monitoring
Tobacco Use. Chronic Dis Can 1994;15(3):105–10.
100
Chronic Diseases in Canada
Workshop participants
Richard Gallagher,* Jason Rivers* (BC Cancer Agency);
Loraine Marrett (Cancer Care Ontario and University of
Toronto); Angus Fergusson (Environment Canada); Christina
Mills,* Pascale Reinhardt-Poulin, Yvon Deslauriers (Health
Canada); Fredrick Ashbury—facilitator (PICEPS Consultant
Inc.); Louise DeGuire (Montreal Department of Public
Health); Sharon Campbell* (National Cancer Institute of
Canada); Cynthia Jorgensen (US Centers for Disease Control
and Prevention); Chris Lovato,* Jean Shoveller,* Larry
Peters (University of British Columbia); Louise Potvin
(University of Montreal); Irving Rootman (University of
Toronto); Anthony Cullen (University of Waterloo)
* Member of workshop planning committee O
Vol 20, No 2
Book Review
Injury Prevention: An International Perspective
Epidemiology, Surveillance, and Policy
By Peter Barss, Gordon S Smith, Susan P Baker
and Dinesh Mohan
New York: Oxford University Press, 1998; xii + 375 pp;
ISBN 0-19-511982-7; $85.50 (CAN)
This book is a valuable resource for those involved in
injury surveillance and prevention. It is international in
scope and explores injury issues not only of developed
countries, but also from the perspective of less developed
countries and of indigenous peoples around the world.
The ideas and data presented combined with
discussion of complex causal factors will engage the
interest of public health professionals and other experts
involved in injury prevention. Yet the uncomplicated
manner in which the data are presented and the flowing,
easy-reading style will also make it useful for anyone
who is interested in injury prevention.
The book offers international data for people of all
ages, including both intentional and unintentional
injuries. It is well organized, starting with an overview of
the epidemiology of injury, profiling international data
on overall injury mortality and morbidity, and then
discussing important categories of injury such as traffic
injuries, drownings, falls, burns, poisonings,
occupational injuries, homicides and suicides. Also
included are sections on the costs of injury, treatment
and rehabilitation.
The authors have gathered an impressive selection of
international injury data. They try to present the data by
age group and sex for a range of developing countries
from all parts of the world and include comparison data
from a few developed countries. Unfortunately,
Canadian data are not routinely reported in this way
although Canadian examples are frequently used in
discussion, particularly illustrations from aboriginal
communities.
sensitivity to the diverse contributions of population
structure, environment, culture, social factors and
political situation in patterns of injury.
The book includes many useful tables, most of which
provide injury data. However, graphs and figures are
used sparingly and the only illustration is the photograph
on the cover. The pages of text are presented with wide
margins into which sidebars are strategically placed.
These sidebars feature short summaries, specific
examples and interesting quotations.
The following 1949 quotation from John E Gordon is
given at the opening of the first chapter. I think it
provides an excellent introduction to the book and might
also have served as the conclusion.
The newer concept of prevention, as it developed,
was applied almost wholly to disease, to the sick.
The injured were largely forgotten...
Overall rating:
Very good to Excellent
Strengths:
International perspective, good data from
diverse sources and in-depth discussion and
interpretation
Weaknesses:
Limited number of graphs, charts and
illustrations
Audience:
International medical and public health
professionals, government regulators and
policy makers, and others interested or
involved in injury surveillance and prevention
Margaret Herbert
Bureau of Reproductive and Child Health
Laboratory Centre for Disease Control
Health Canada, Tunney’s Pasture
Address Locator: 0601E2
Ottawa, Ontario K1A 0L2
Not all of the data shown are current, many are more
than a decade old. This situation reflects the unfortunate
lack of sound international injury surveillance rather than
any inadequacy in data collection for the book. The
authors compare injury rates across countries and
provide detailed interpretation of the data as well as
insightful discussion of exposure to hazards, complex
causal factors and approaches to prevention. The
discussion reveals remarkable understanding and
1999
101
August 31–Sept 4, 1999
Florence, Italy
“Epidemiology for Sustainable Health”
15th International Scientific Meeting of the
International Epidemiological Association
Organizing Secretariat
IEA Florence ‘99
c/o SINEDRION
Via G. Marconi, 27
50131 Firenze, Italy
Tel: 39-55-570502
Fax: 39-55-575679
E-mail: [email protected]
<http://iea99ds.unifi.it>
October 1–3, 1999
Ottawa, Ontario
“Building and Enriching Partnerships in the
Management of COPD”
Conference presented by The Canadian Chronic
Obstructive Pulmonary Disease (COPD)
Alliance
The Canadian Lung Association
National Office
508 – 1900 City Park Drive
Gloucester, Ontario K1J 1A3
Tel: (613) 747-6776
Fax: (613) 747-7430
E-mail: [email protected]
<www.lung.ca/CCA/conference>
October 1–3, 1999
Toronto, Ontario
“Closing the Loop: Evidence into Health
Practice, Organization and Policy”
3rd International Conference the Scientific Basis
of Health Services
Sponsors: Canadian Foundation for Health
Services Research, Health Canada and
Medical Research Council of Canada
3rd Intl Conference on the Scientific
Basis of Health Services
c/o Alysone Will, CMP
The Paragon Conference & Event Group
Inc.
704 – 205 Richmond Street West
Toronto, Ontario M5V 1V3
Tel: (416) 979-1300
Fax: (416) 979-1819
E-mail:
[email protected]
<www.paragon-conferences.on.ca/
health99.html>
October 18–20, 1999
Chilton, Oxfordshire
United Kingdom
International Workshop on UV Exposure,
Measurement and Protection
Sponsors: National Radiological Protection
Board (NRPB), World Health Organization and
International Commission on Non-Ionizing
Radiation Protection
Dr Colin Driscoll
NRPB (UV Workshop)
Chilton, Didcot, OX11 0RQ
United Kingdom
Tel: 44-1235-822724
Fax: 44-1235-831600
E-mail: [email protected]
<http://www.nrpb.org.uk/
WHO-uv.htm>
November 11–13, 1999
Halifax, Nova Scotia
Fourth National Conference on Asthma and
Education (ASED 4)
Presented by the Canadian Network for Asthma
Care (CNAC)
A Les McDonald
Executive Director, CNAC
1607 – 6 Forest Laneway
North York, Ontario M2N 5X9
Tel: (416) 224-9221
Fax: (416) 224-9220
E-mail: [email protected]
<www.cnac.net>
102
Chronic Diseases in Canada
Vol 20, No 2
November 15–17, 1999
Toronto, Ontario
“Celebrating Our Past, Building Our Future”
50th Annual Ontario Public Health Association
Conference
Co-hosted by the Centre for Health Promotion
and Toronto Public Health
OPHA Conference Secretary
Toronto Public Health
North York Civic Centre
North York, Ontario M2N 5V7
Tel: (416) 395-7653
Fax: (416) 395-7691
<http://www.web.net/opha/>
November 30–Dec 2, 1999
Dallas, Texas
USA
“Prevention Successes 2000: Better Health
for All”
14th National Conference on Chronic Disease
Prevention and Control
Sponsors: Centers for Disease Control and
Prevention, Association of State and Territorial
Chronic Disease Program Directors and
American Heart Association
Estella Lazenby
The KEVRIC Company, Inc.
610 – 8401 Colesville Road
Silver Spring, MD
USA 20910
Tel: (301) 588-6000
Fax: (301) 588-2106
E-mail: [email protected]
<www.cdc.gov/nccdphp>
March 13–16, 2000
Quebec City, Quebec
“Health and the Quality of Life: Our Municipalities
in an Era of Globalization”
3rd Conference of Local Health Authorities of the
Americas
Organized by l’Institut national de santé publique
du Québec and the WHO Collaborating Centre
for the Development of Healthy Cities and
Villages
Secrétariat du 3e Congrès des
responsables locaux de santé des
Amériques
938, rue Saint-Maurice
Montréal (Québec) H3C 1L7
Tel: (514) 395-1808
Fax: (514) 395-1801
E-mail: [email protected]
<http://www.msss.gouv.qc.ca/
congres_quebec>
May 7–10, 2000
Victoria, British Columbia
“Science and Policy in Action”
First International Conference on Women, Heart
Disease and Stroke
Heart and Stroke Foundation, American Heart
Association, Health Canada and Centers for
Disease Control and Prevention are providing
early leadership
Taylor & Associates
18 – 5370 Canotek Road
Gloucester, Ontario K1J 9E8
Tel: (613) 747-0262
Fax: (613) 745-1846
E-mail: [email protected]
May 28–30, 2000
Ottawa, Ontario
“Charting the Course for Literacy and Health in
the New Millennium”
First Canadian Conference on Literacy and
Health
Organized by the Canadian Public Health
Association’s (CPHA) National Literacy and
Health Program
CPHA Conference Department
400 – 1565 Carling Avenue
Ottawa, Ontario K1Z 8R1
Tel: (613) 725-3769
Fax: (613) 725-9826
E-mail: [email protected]
<www.nald.ca/nlhp.htm>
August 23–27, 2000
Victoria, British Columbia
ITCH 2000: “From Potential to Practice”
International Conference on Information
Technology in Community Health
Call for abstracts and student poster contest
deadline: December 15, 1999
ITCH 2000
c/o School of Health Information
Science
University of Victoria
PO Box 3050, STN CSC
Victoria, BC V8W 3P5
Tel: (250) 721-8576
Fax: (250) 472-4751
E-mail: [email protected]
<http://itch.uvic.ca>
1999
103
More 1998 Peer Reviewers
Our list of people to thank for doing peer review
for us in 1998 (published in Volume 20, No 1)
should have included two more names.
Gail Eyssen
Ken Johnson
Announcement
New journal now available
Cancer Strategy
ISSN: 1464-1828 / Quarterly
Editor-in-Chief: Karol Sikora, Chief, Cancer
Programme, World Health Organization,
International Agency for Research on
Cancer
Free online sample copy available at
<http://www.stockton-press.co.uk/cs>
104
Chronic Diseases in Canada
Vol 20, No 2
CDIC: Information for Authors
Chronic Diseases in Canada (CDIC) is a peer-reviewed scientific
journal published four times a year. Contributions are welcomed
from outside of Health Canada as well as from within this federal
department. The journal's focus is the prevention and control of
non-communicable diseases and injuries in Canada. This may
include research from such fields as epidemiology,
public/community health, biostatistics, behavioural sciences and
health services. CDIC endeavours to foster communication among
public health practitioners, chronic disease epidemiologists and
researchers, health policy planners and health educators.
Submissions are selected based on scientific quality, public health
relevance, clarity, conciseness and technical accuracy. Although
CDIC is a Health Canada publication, authors retain responsibility
for the contents of their papers, and opinions expressed are not
necessarily those of the CDIC Editorial Committee or of Health
Canada.
Each submission must have a covering letter signed by all authors
that identifies the corresponding author (including fax number)
and states that all authors have seen and approved the final
manuscript and have met the authorship criteria of the Uniform
Requirements. The covering letter should also include a full
statement regarding any prior or duplicate publication or
submission for publication. Written permission from anyone
mentioned by name in the acknowledgements should appear at
this time. Suggestions for appropriate peer reviewers are
appreciated as well.
Feature Articles
If a manuscript is accepted for publication, send the final hardcopy
version with the accompanying text file in WordPerfect or ASCII,
in IBM-compatible format, specifying the software version.
Most feature articles are limited to 3500 words of text in the form
of original research, surveillance reports, meta-analyses,
methodological papers or literature reviews. The maximum length
for Short Reports is 1500 words, and Position Papers should not
exceed 3000 words.
Under normal circumstances, two other types of feature articles
(both 3000 words maximum) will be considered as submissions
only from authors within Health Canada: Status Reports
describing ongoing national programs, studies or information
systems of interest to chronic disease researchers and public health
practitioners; and Workshop/Conference Reports of relevant
workshops, etc. organized or sponsored by Health Canada.
Authors outside of Health Canada may submit reports for our
Cross-country Forum (3000 words maximum) to exchange
information and insights about the prevention and control of
chronic diseases and injuries from research or surveillance
findings, programs under development or program evaluations.
Additional Article Types
Letters to the Editor (500 words maximum) commenting on
articles recently published in CDIC will be considered for
publication. Book/Software Reviews (1300 words maximum) are
usually solicited by the editors. In addition, the editors
occasionally solicit Guest Editorials.
Submitting Manuscripts
Submit manuscripts to the Editor-in-Chief, Chronic Diseases in
Canada, Laboratory Centre for Disease Control, Health Canada,
Tunney's Pasture, CDIC Address Locator: 0602C3, Ottawa,
Ontario K1A 0L2.
Since Chronic Diseases in Canada adheres in general (section on
illustrations not applicable) to the “Uniform Requirements for
Manuscripts Submitted to Biomedical Journals” as approved
by the International Committee of Medical Journal Editors,
authors should refer to the Canadian Medical Association Journal
1997 Jan 15; 156(2): 270–7 for complete details (or at
www.cma.ca/publications/mwc/uniform.htm).
Manuscripts may be submitted in either English or French and will
be published in both languages, if accepted. Submit four complete
printed copies of a manuscript, double-spaced, on standard-sized
paper with one-inch margins. Each section (i.e. title page, abstract
and key words, text, acknowledgements, references, tables and
figures) should begin on a separate, numbered page.
Abstract and Key Words
An unstructured abstract not exceeding 150 words (100 words
only for Short Reports) must accompany each manuscript with
three to eight key words noted below, preferably from the Medical
Subject Headings (MeSH) of Index Medicus.
Tables and Figures
Tables and figures should be as self-explanatory and succinct as
possible. They should not simply duplicate the text, but should
illuminate and supplement it, and they should not be too
numerous. Place them on separate pages after the references,
numbered in the order that they are mentioned in the text.
Provide explanatory material for tables in footnotes, identifying
the table footnotes by lower-case superscript letters in alphabetical
order.
Figures must be limited to graphs or flow charts/templates; we are
unable to publish photographic illustrations at this time. Specify
the software used (preferably Harvard Graphics) and supply raw
data (in hardcopy form) for all graphs. Do not import figures into
the text of the manuscript.
Authors must obtain written permission from the copyright holder
to reproduce or adapt any tables or figures that have been
published previously.
References
References should follow the Vancouver style, numbered
consecutively in the order that they first appear in the text
(identified by numbers in superscript or within parentheses) and
arranged numerically in the reference list. References cited only in
tables or figures should be numbered as above according to the
first mention of the particular table/figure in the text. Remove any
endnote/footnote word-processing feature used to generate a
reference list.
Authors are responsible for verifying the accuracy of references.
The use of references to unpublished observations/data or personal
communications is discouraged; if used, do not include with
numbered references but in the text in parentheses and obtain
permission for these citations.
Chronic Diseases in Canada
a publication of the
Laboratory Centre for Disease Control
Health Protection Branch
Health Canada
Editor-in-Chief ............................ Lori Anderson
Scientific Editor ....................... Christina J Mills
Associate Scientific Editor ............. Gerry B Hill
Associate Scientific Editor ........ Stephen B Hotz
Desktop Publisher ........................ Holly Dopson
CDIC Editorial Committee
Donald T Wigle, Committee Chair
Laboratory Centre for Disease Control
Health Canada
Jean-François Boivin
McGill University
Jacques Brisson
Université Laval
Neil E Collishaw
World Health Organization
James A Hanley
McGill University
Chronic Diseases in Canada (CDIC) is a quarterly scientific journal
focusing on current evidence relevant to the control and prevention of
chronic (i.e. non-communicable) diseases and injuries in Canada.
Feature articles may include research from such fields as epidemiology,
public/community health, biostatistics, behavioural sciences and health
services. Scientific articles are peer reviewed. The journal publishes a
unique blend of public and private sector authors, with information for
authors in every issue. Subscription is available free upon request.
Authors retain responsibility for the contents of their papers, and
opinions expressed are not necessarily those of the CDIC Editorial
Committee or of Health Canada.
When submitting change of address,
please enclose your old address label.
Mailing Address:
Chronic Diseases in Canada
Laboratory Centre for Disease Control
Health Canada, Tunney's Pasture
Address Locator: 0602C3
Ottawa, Ontario K1A 0L2
Telephone:
Editor-in-Chief
Scientific Editor
Circulation
Fax
Clyde Hertzman
University of British Columbia
C Ineke Neutel
Therapeutic Products Directorate
Health Canada
Kathryn Wilkins
Health Statistics Division
Statistics Canada
(613) 957-1767
(613) 957-2624
(613) 941-1291
(613) 952-7009
Indexed in Index Medicus/MEDLINE and PAIS
(Public Affairs Information Service)
This publication can also be accessed electronically on the World Wide Web site
of the Laboratory Centre for Disease Control at <http://www.hc-sc.gc.ca/hpb/lcdc>.
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