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. 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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 12. 1. 2. 3. 4. 5. 6. 7. 8. 9. 1999 Newcombe HB, Kennedy JM, Axford SJ. Automatic linkage of vital records. Science 1959;130:954–9. Howe GR, Lindsay J. A generalized iterative record linkage computer system for use in medical follow-up studies. Comput Biomed Res 1981;14:327–40. Newcombe HB. Handbook of record linkage: methods for health and statistical studies, administration, and business. Oxford, England: Oxford University Press, 1988. Miller AB, Howe GR, Sherman GJ. Mortality from breast cancer after irradiation during fluoroscopic examinations in patients being treated for tuberculosis. N Engl J Med 1989;321:1285–9. Herderson J, Goldacre MJ, Graveney MJ, Simmons HM. Use of medical record linkage to study readmission rates. Br Med J 1989;299:709–13. Van Den Brandt PA, Schouten LJ, Goldbohm RA, Dorant E, Hunen PMH. Development of a record linkage protocol for use in the Dutch Cancer Registry for epidemiological research. Int J Epidemiol 1990;19:553–8. Roos LL, Wajda A. Record linkage strategies. Part I: estimating information and evaluating approaches. Meth Inform Med 1991;30:117–23. Goldberg MS, Carpenter M, Thériault G, Fair M. The accuracy of ascertaining vital status in a historical cohort study of synthetic textiles workers using computerized record linkage to the Canadian Mortality Data Base. Can J Public Health 1993;84:201–4. Nash JQ, Chandrakumar M, Farrington CP, Williamson S, Miller E. Feasibility study for identifying adverse 13. 14. 15. 16. 17. 18. 19. events attributable to vaccination by record linkage. Epidemiol Infect 1995;114:475–80. The West of Scotland Coronary Prevention Study Group. Computerised record linkage: compared with traditional patient follow-up methods in clinical trials and illustrated in a prospective epidemiological study. J Clin Epidemiol 1995;48:1441–52. Jamieson E, Roberts J, Browne G. The feasibility and accuracy of anonymized record linkage to estimate shared clientele among three health and social service agencies. Meth Inform Med 1995;34:371–7. Howe GR. Lung cancer mortality between 1950 and 1987 after exposure to fractionated moderate-dose-rate ionizing radiation in the Canadian fluoroscopy cohort study and a comparison with lung cancer mortality in the atomic bomb survivors study. Radiat Res 1995;142:295–304. Howe GR, McLaughlin J. Breast cancer mortality between 1950 and 1987 after exposure to fractionated moderate-dose-rate ionizing radiation in the Canadian fluoroscopy cohort study and a comparison with breast cancer mortality in the atomic bomb survivors study. Radiat Res 1996;145:694–707. Herrchen B, Gould JB, Nesbitt TS. Vital statistics linked birth/infant death and hospital discharge record linkage for epidemiological studies. Comput Biomed Res 1997;30:290–305. 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 method for conducting emergencey medical services cohort studies using existing data. Acad Emerg Med 1997;4:1087–95. Howe GR. Use of computerized record linkage in cohort studies. Epidemiol Rev 1998;20:112–21. Wen SW, Liu S, Fowler D. Trends and variations in neonatal length of in-hospital stay in Canada. Can J Public Health 1998;89:115–9. Wen SW, Liu S, Marcoux S, Fowler D. Uses and limitations of routine hospital admission/separation records for perinatal surveillance. Chronic Dis Can 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. 94 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. 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