Am. J. Hum. Genet. 49:1014-1024, 1991
Epidemiology of Down Syndrome in South Australia, 1960-89 Alan J. Staples,* Grant R. Sutherland,t Eric A. Haan,* and Susan Clisbyt Departments of Medical Genetics and Epidemiology and tCytogenetics and Molecular Genetics, Adelaide Children's Hospital, and tIntellectually Disabled Services Council, Incorporated, North Adelaide
Summary During 1960-89 687 Down syndrome live births and 46 Down syndrome pregnancy terminations were identified in South Australia. Ascertainment was estimated to be virtually complete. The sex distribution of Down syndrome live births was found to be statistically different from the non-Down syndrome live-birth sex distribution (P < .01). Smoothed maternal age-specific incidence was derived using both maternal age calculated to the nearest month and a discontinuous-slope regression model. The incidence of Down syndrome at birth for the study period was estimated to be 1.186 Down syndrome births/1,000 live births. Annual population incidence was shown to be correlated with trends in the maternal age distribution of confinements. If current trends in the maternal age distribution of confinements continue, the population incidence of Down syndrome in South Australia is predicted to exceed 1.5 Down syndrome births/1,000 live births during the 1990-94 quinquennium. Introduction
The last comprehensive epidemiological study of Down syndrome in South Australia was undertaken during 1978 and covered the period 1955-77 (Sutherland et al. 1979). Maternal age-related risk figures were produced, and these have since been used to counsel older mothers about their risk of giving birth to a child with Down syndrome. In South Australia, chorion villus sampling or amniocentesis is available to women aged 35 years or older, and approximately 50O in this age group choose one of these definitive tests for Down syndrome. More recently, methods have been devised to screen the entire population of pregnant women to determine each woman's individual risk of carrying a Down syndrome child (Wald et al. 1988). Screening is performed by measuring selected analytes-e.g., a-fetoprotein, human chorionic gonadotrophin, and unconjugated estriol - in maternal serum in the second Received December 10, 1990; final revision received July 2, 1991.
Address for correspondence and reprints: Grant R. Sutherland, Department of Cytogenetics and Molecular Genetics, Adelaide Children's Hospital, North Adelaide, South Australia 5006. i 1991 by The American Society of Human Genetics. All rights reserved. 0002-9297/91 /4905-0012$02.00
1014
trimester. These results are combined with maternal age to produce a risk figure for the birth of a Down syndrome child. Women can choose to proceed to amniocentesis if they are shown to be at increased risk. Screening the entire pregnant population in this way could detect over 60% of Down syndrome pregnancies (Wald et al. 1988). We have undertaken a detailed review of the birth prevalence of Down syndrome in South Australia so that precise age-related risk figures can be incorporated into our proposed algorithm for a local maternal serum screening program. The data presented in the present study are unique in that they are from a defined population, the state of South Australia, for which there is very complete demographic information (Australian Bureau of Statistics 1989), and they represent virtual total ascertainment over a 30-year period. In 1989 South Australia had a population of 1.425 million. Methods
Sources of Ascertainment Data collected in the previous Down syndrome study (Sutherland et al. 1979) were used to establish the South Australian Down Syndrome Register (estab-
Down Syndrome in South Australia lished in 1978). The only two cytogenetic laboratories in South Australia provide information concerning cytogenetically diagnosed (prenatal and postnatal) cases of Down syndrome to this register. The Intellectually Disabled Services Council is a state government body which provides and coordinates all services for the intellectually handicapped in South Australia (established in 1982). All Down syndrome individuals who received such services were ascertained. The South Australian Birth Defects Register (established in 1986) receives notification of Down syndrome live births, stillbirths, and pregnancy terminations from a variety of sources, including the South Australian Health Commission's Pregnancy Outcome Statistics Unit (to which all terminations of pregnancy are notified; established in 1977), the two cytogenetic laboratories, public and private hospitals, and private medical practitioners. Data from all these sources were pooled and validated. The rate of multiple ascertainment was not recorded but was very high. Down syndrome parental and child date-of-birth data were confirmed by the South Australian Register of Births, Deaths and Marriages. Population data for South Australia, detailing the total number of live births (by maternal age in years) for each year of the study period and parental age data, were supplied by the Australian Bureau of Statistics. Data Analysis
Data analysis had three objectives: (1) to identify and enumerate all Down syndrome live births and prenatally diagnosed terminations during the period of study, (2) to determine whether the maternal agespecific incidence of Down syndrome had changed with time during the period of study, and (3) to derive a regression model describing maternal age-specific incidence of Down syndrome in South Australia (a model which may be incorporated into an antenatal serum screening program). Data analysis involved the production of tableau information to describe the epidemiology of Down syndrome, tests of comparison to detect changes with time and to validate statistical inferences, and regression analysis to estimate maternal age-specific incidence. Ascertainment of Down syndrome live births is likely to have been complete during the past 20 years of the study. A small number of Down syndrome live births may not have been ascertained during the first 10 years of the study period. Ascertainment of terminated Down syndrome pregnancies is known to be complete, since only two laboratories are involved in
1015
the process. Since the ascertainment of Down syndrome live births was virtually complete, no adjustment for any ascertainment bias was made. Adjustments Applied to Terminated Down Syndrome Pregnancies Approximately 30% of Down syndrome fetuses are lost after 16 wk of pregnancy (Hook 1983), so each terminated Down syndrome pregnancy was assumed to be equivalent to 0.7 live births. These estimated live births were assumed to be born at 38 wk gestation. If the gestational age at termination of pregnancy was unknown (nine cases), the mean gestational age for Down syndrome at time of termination was assumed (18 wk), and 20 wk was added to the termination date, to yield a date of birth estimate. The maternal age at estimated date of delivery for a terminated Down syndrome pregnancy was similarly adjusted. These births were added to both the population livebirth data and the Down syndrome live-birth data for the corresponding maternal age and probable year of birth. Adjustments Applied to Down Syndrome Maternal Ages
Maternal date of birth was unknown for 10 liveborn and six terminated cases of Down syndrome. However, maternal year of birth or maternal age in completed years was available for all but one Down syndrome live birth. For these 15 cases, the maternal age was adjusted either by adding 6 mo if maternal age in completed years was known or by assuming that the mother was born midyear if the year of birth was identified. The case for which maternal age could not be identified was omitted from the calculation of summary statistics pertaining to maternal age and maternal age-specific incidence but was included in all other analyses. Tests of Comparison All contingency-table tests were computed using the X2 with Yates's correction for continuity. Regression Analysis Various regression-model functional forms have been used to describe the maternal age-specific incidence of Down syndrome (Lamson and Hook 1981). These include the constant-plus-exponential model (Huether et al. 1981), power model (Penrose and Smith 1966, pp. 162-171) (X'), constant-plusPoisson model (Penrose and Smith 1966, pp. 162171), polynomial models (Hook and Chambers
Staples et al. (I)
1016
1977), and the discontinuous-slope model (Hook and Chambers 1977; Hook and Fabia 1978; Hook and Lindsjo 1978). No single functional form has been shown to be consistently superior in describing the maternal age-specific incidence of Down syndrome. In addition to the models referred to above, other more complex models have been used successfully. Several functional forms were evaluated in the present study. The model sought was one which was mathematically simple and consistent with the South Australian data. The preferred functional form was selected on the criteria of (a) a comparison of adjusted R2 statistics (Neter and Wasserman 1974, pp. 228229, 376-378) when regression variables for any two models were of the same numerical form and (b) a comparison of the log-likelihood statistic (Pollard 1977,pp.18-25,211-212)G2(G2= 2log[likelihood ratio]) when regression variables for any two models were of different numerical form. Regression analysis of age-specific incidence was calculated by regressing the maternal age at time of birth (or, in the case of a terminated pregnancy, estimated date of delivery), to the nearest month, against the maternal age proportion of Down syndrome live births. The maternal age proportion of Down syndrome live births was calculated by dividing the number of Down syndrome live births by the total number of live births for each maternal age (in completed years). This analytical approach provides a weighted least-squares regression and minimizes any maternal age bias present within a single maternal year-of-age interval. The constant plus exponential, constant plus Poisson, and power models were discarded, as they provided a poor fit to the data at ages under 32 years and were inferior, log-likelihood criteria, to alternative models on adjusted R2 and G2. Regression analysis implied that no pure model was preferred over the whole maternal age range. A linear discontinuousslope model provided the best fit for maternal ages under 32 years, and a quadratic polynomial model (in natural logarithms) provided the best fit for maternal ages over 32 years. The residuals of the discontinuous-slope model were not randomly distributed for maternal ages exceeding 32 years. These results suggested that one of the following should be attempted: either (a) a hybrid model comprising both a linear component for maternal ages under 32 years and a quadratic component for maternal ages above 32 years or (b) the switching of regression parameters at about age 30. A comparison of adjusted R2 statistics, G2 log-likelihood statistics, residual analysis, and con-
sideration of both parameter estimation efficiency and the intended application of the model led to the selection of a linear discontinuous-slope model with switching of parameters at age 30 years: In y = bo + bix, + b2x2 and y = exp (bo + bix, + b2 X2), where y = incidence/ 1,000 live births, x1 = maternal age minus 30 (if age 30 years), and X2 = maternal age minus 30 (if age >30 years) and 0 (if age
Epidemiology of Down Syndrome in South Australia, 1960-89 Alan J. Staples,* Grant R. Sutherland,t Eric A. Haan,* and Susan Clisbyt Departments of Medical Genetics and Epidemiology and tCytogenetics and Molecular Genetics, Adelaide Children's Hospital, and tIntellectually Disabled Services Council, Incorporated, North Adelaide
Summary During 1960-89 687 Down syndrome live births and 46 Down syndrome pregnancy terminations were identified in South Australia. Ascertainment was estimated to be virtually complete. The sex distribution of Down syndrome live births was found to be statistically different from the non-Down syndrome live-birth sex distribution (P < .01). Smoothed maternal age-specific incidence was derived using both maternal age calculated to the nearest month and a discontinuous-slope regression model. The incidence of Down syndrome at birth for the study period was estimated to be 1.186 Down syndrome births/1,000 live births. Annual population incidence was shown to be correlated with trends in the maternal age distribution of confinements. If current trends in the maternal age distribution of confinements continue, the population incidence of Down syndrome in South Australia is predicted to exceed 1.5 Down syndrome births/1,000 live births during the 1990-94 quinquennium. Introduction
The last comprehensive epidemiological study of Down syndrome in South Australia was undertaken during 1978 and covered the period 1955-77 (Sutherland et al. 1979). Maternal age-related risk figures were produced, and these have since been used to counsel older mothers about their risk of giving birth to a child with Down syndrome. In South Australia, chorion villus sampling or amniocentesis is available to women aged 35 years or older, and approximately 50O in this age group choose one of these definitive tests for Down syndrome. More recently, methods have been devised to screen the entire population of pregnant women to determine each woman's individual risk of carrying a Down syndrome child (Wald et al. 1988). Screening is performed by measuring selected analytes-e.g., a-fetoprotein, human chorionic gonadotrophin, and unconjugated estriol - in maternal serum in the second Received December 10, 1990; final revision received July 2, 1991.
Address for correspondence and reprints: Grant R. Sutherland, Department of Cytogenetics and Molecular Genetics, Adelaide Children's Hospital, North Adelaide, South Australia 5006. i 1991 by The American Society of Human Genetics. All rights reserved. 0002-9297/91 /4905-0012$02.00
1014
trimester. These results are combined with maternal age to produce a risk figure for the birth of a Down syndrome child. Women can choose to proceed to amniocentesis if they are shown to be at increased risk. Screening the entire pregnant population in this way could detect over 60% of Down syndrome pregnancies (Wald et al. 1988). We have undertaken a detailed review of the birth prevalence of Down syndrome in South Australia so that precise age-related risk figures can be incorporated into our proposed algorithm for a local maternal serum screening program. The data presented in the present study are unique in that they are from a defined population, the state of South Australia, for which there is very complete demographic information (Australian Bureau of Statistics 1989), and they represent virtual total ascertainment over a 30-year period. In 1989 South Australia had a population of 1.425 million. Methods
Sources of Ascertainment Data collected in the previous Down syndrome study (Sutherland et al. 1979) were used to establish the South Australian Down Syndrome Register (estab-
Down Syndrome in South Australia lished in 1978). The only two cytogenetic laboratories in South Australia provide information concerning cytogenetically diagnosed (prenatal and postnatal) cases of Down syndrome to this register. The Intellectually Disabled Services Council is a state government body which provides and coordinates all services for the intellectually handicapped in South Australia (established in 1982). All Down syndrome individuals who received such services were ascertained. The South Australian Birth Defects Register (established in 1986) receives notification of Down syndrome live births, stillbirths, and pregnancy terminations from a variety of sources, including the South Australian Health Commission's Pregnancy Outcome Statistics Unit (to which all terminations of pregnancy are notified; established in 1977), the two cytogenetic laboratories, public and private hospitals, and private medical practitioners. Data from all these sources were pooled and validated. The rate of multiple ascertainment was not recorded but was very high. Down syndrome parental and child date-of-birth data were confirmed by the South Australian Register of Births, Deaths and Marriages. Population data for South Australia, detailing the total number of live births (by maternal age in years) for each year of the study period and parental age data, were supplied by the Australian Bureau of Statistics. Data Analysis
Data analysis had three objectives: (1) to identify and enumerate all Down syndrome live births and prenatally diagnosed terminations during the period of study, (2) to determine whether the maternal agespecific incidence of Down syndrome had changed with time during the period of study, and (3) to derive a regression model describing maternal age-specific incidence of Down syndrome in South Australia (a model which may be incorporated into an antenatal serum screening program). Data analysis involved the production of tableau information to describe the epidemiology of Down syndrome, tests of comparison to detect changes with time and to validate statistical inferences, and regression analysis to estimate maternal age-specific incidence. Ascertainment of Down syndrome live births is likely to have been complete during the past 20 years of the study. A small number of Down syndrome live births may not have been ascertained during the first 10 years of the study period. Ascertainment of terminated Down syndrome pregnancies is known to be complete, since only two laboratories are involved in
1015
the process. Since the ascertainment of Down syndrome live births was virtually complete, no adjustment for any ascertainment bias was made. Adjustments Applied to Terminated Down Syndrome Pregnancies Approximately 30% of Down syndrome fetuses are lost after 16 wk of pregnancy (Hook 1983), so each terminated Down syndrome pregnancy was assumed to be equivalent to 0.7 live births. These estimated live births were assumed to be born at 38 wk gestation. If the gestational age at termination of pregnancy was unknown (nine cases), the mean gestational age for Down syndrome at time of termination was assumed (18 wk), and 20 wk was added to the termination date, to yield a date of birth estimate. The maternal age at estimated date of delivery for a terminated Down syndrome pregnancy was similarly adjusted. These births were added to both the population livebirth data and the Down syndrome live-birth data for the corresponding maternal age and probable year of birth. Adjustments Applied to Down Syndrome Maternal Ages
Maternal date of birth was unknown for 10 liveborn and six terminated cases of Down syndrome. However, maternal year of birth or maternal age in completed years was available for all but one Down syndrome live birth. For these 15 cases, the maternal age was adjusted either by adding 6 mo if maternal age in completed years was known or by assuming that the mother was born midyear if the year of birth was identified. The case for which maternal age could not be identified was omitted from the calculation of summary statistics pertaining to maternal age and maternal age-specific incidence but was included in all other analyses. Tests of Comparison All contingency-table tests were computed using the X2 with Yates's correction for continuity. Regression Analysis Various regression-model functional forms have been used to describe the maternal age-specific incidence of Down syndrome (Lamson and Hook 1981). These include the constant-plus-exponential model (Huether et al. 1981), power model (Penrose and Smith 1966, pp. 162-171) (X'), constant-plusPoisson model (Penrose and Smith 1966, pp. 162171), polynomial models (Hook and Chambers
Staples et al. (I)
1016
1977), and the discontinuous-slope model (Hook and Chambers 1977; Hook and Fabia 1978; Hook and Lindsjo 1978). No single functional form has been shown to be consistently superior in describing the maternal age-specific incidence of Down syndrome. In addition to the models referred to above, other more complex models have been used successfully. Several functional forms were evaluated in the present study. The model sought was one which was mathematically simple and consistent with the South Australian data. The preferred functional form was selected on the criteria of (a) a comparison of adjusted R2 statistics (Neter and Wasserman 1974, pp. 228229, 376-378) when regression variables for any two models were of the same numerical form and (b) a comparison of the log-likelihood statistic (Pollard 1977,pp.18-25,211-212)G2(G2= 2log[likelihood ratio]) when regression variables for any two models were of different numerical form. Regression analysis of age-specific incidence was calculated by regressing the maternal age at time of birth (or, in the case of a terminated pregnancy, estimated date of delivery), to the nearest month, against the maternal age proportion of Down syndrome live births. The maternal age proportion of Down syndrome live births was calculated by dividing the number of Down syndrome live births by the total number of live births for each maternal age (in completed years). This analytical approach provides a weighted least-squares regression and minimizes any maternal age bias present within a single maternal year-of-age interval. The constant plus exponential, constant plus Poisson, and power models were discarded, as they provided a poor fit to the data at ages under 32 years and were inferior, log-likelihood criteria, to alternative models on adjusted R2 and G2. Regression analysis implied that no pure model was preferred over the whole maternal age range. A linear discontinuousslope model provided the best fit for maternal ages under 32 years, and a quadratic polynomial model (in natural logarithms) provided the best fit for maternal ages over 32 years. The residuals of the discontinuous-slope model were not randomly distributed for maternal ages exceeding 32 years. These results suggested that one of the following should be attempted: either (a) a hybrid model comprising both a linear component for maternal ages under 32 years and a quadratic component for maternal ages above 32 years or (b) the switching of regression parameters at about age 30. A comparison of adjusted R2 statistics, G2 log-likelihood statistics, residual analysis, and con-
sideration of both parameter estimation efficiency and the intended application of the model led to the selection of a linear discontinuous-slope model with switching of parameters at age 30 years: In y = bo + bix, + b2x2 and y = exp (bo + bix, + b2 X2), where y = incidence/ 1,000 live births, x1 = maternal age minus 30 (if age 30 years), and X2 = maternal age minus 30 (if age >30 years) and 0 (if age
