Parity and the Risk of Down’s Syndrome

V. Paul Doria-Rose1,2 , Han S. Kim1,2, Elizabeth T. J. Augustine1 and Karen L. Edwards1

1 Department of Epidemiology, School of Public Health and Community Medicine, University of Washington, Seattle, WA.
2 Division of Public Health Sciences, Fred Hutchinson Cancer Research Center, Seattle, WA.

Received for publication October 15, 2002; accepted for publication March 3, 2003.


    ABSTRACT
 TOP
 ABSTRACT
 INTRODUCTION
 MATERIALS AND METHODS
 RESULTS
 DISCUSSION
 REFERENCES
 
Interpretation of studies that have examined parity as a risk factor for Down’s syndrome has been hindered by inadequate control for maternal age and/or failure to account for the differential use of prenatal diagnosis and pregnancy termination between low-parity and high-parity women. In this case-control study, the authors used exact matching on maternal age–minimize confounding and evaluated the potential impact of differential termination. A total of 898 cases of Down’s syndrome and 4,488 controls were identified using Washington State birth certificates from 1984–1998. There was a trend towards increasing risk of Down’s syndrome with increasing parity in both younger (age <35 years) and older (age >=35 years) mothers. Restriction to women with no indication of amniocentesis (for whom differential termination is unlikely) resulted in a blunting of the odds ratios; however, a trend for parity remained. After restriction, odds ratios were as high as 1.65 (95% confidence interval: 1.13, 2.40) in younger women with a parity of three (compared with a parity of zero) and 2.41 (95% confidence interval: 1.41, 4.12) in older women with a parity of four or more. Although the odds ratios for older women were probably biased upwards because of underreporting of amniocentesis on birth certificates, these data support an association between parity and Down’s syndrome.

case-control studies; Down syndrome; parity; prenatal diagnosis; prevalence; risk factors

Abbreviations: Abbreviation: BERD, Birth Events Record Database.


    INTRODUCTION
 TOP
 ABSTRACT
 INTRODUCTION
 MATERIALS AND METHODS
 RESULTS
 DISCUSSION
 REFERENCES
 
Editor’s note: An invited commentary on this article appears on page 509, and the authors’ response appears on page 512.

Down’s syndrome is one of the most frequently reported birth defects in the United States, with a prevalence of 9.2 cases per 10,000 livebirths (1). Advanced maternal age is strongly associated with an increased risk of Down’s syndrome; however, other risk factors are less well established. Investigators from several studies have reported a positive association between parity and Down’s syndrome (26), although several other groups did not find an association (79). The interpretation of many of these studies has been hindered by certain methodological issues (10). Because parity is closely correlated with maternal age, and because several early studies examining the relation between parity and Down’s syndrome used broad (5-year) categories in controlling for maternal age (25), it has been suggested that at least part of the apparent effect of parity is due to residual confounding (4, 10). Additionally, there is some evidence that women of higher parity are less likely to undergo prenatal screening for Down’s syndrome by amniocentesis or chorionic villus sampling (6, 8, 1113) and therefore are less likely to choose to terminate a Down’s syndrome pregnancy than women of lower parity. This would result in an excess of Down’s syndrome livebirths among multiparous women, even in the absence of a true biologic association with parity.

The purpose of this study was to examine increasing parity as an independent risk factor for Down’s syndrome, while attempting to minimize the possibility of residual confounding by age, and to evaluate the potential effect of differential pregnancy termination practices between high-parity and low-parity women.


    MATERIALS AND METHODS
 TOP
 ABSTRACT
 INTRODUCTION
 MATERIALS AND METHODS
 RESULTS
 DISCUSSION
 REFERENCES
 
A population-based, matched case-control study was conducted, using singleton birth certificate records from the state of Washington from 1984–1998. Cases were infants with Down’s syndrome who were identified using the birth records themselves, based on check-box indication of Down’s syndrome (14), and/or the Birth Events Record Database (BERD). BERD, which was available for the years 1987–1996, links Washington State hospital inpatient discharge records for mothers and newborns with birth and infant death records, providing information on hospitalizations, prenatal care, and birth outcomes. Cases were identified using BERD as those infants with an International Classification of Diseases, Ninth Revision, code of 758.0 for Down’s syndrome. A total of 898 cases were identified. For the years in which BERD was available, 164 cases (24 percent) were identified through the birth certificate records only, 262 (38 percent) through the BERD database only, and 270 (39 percent) through both BERD and birth certificate records. An additional 202 cases were identified by birth records only in years for which BERD was unavailable.

Five controls were matched to each case on the basis of exact maternal age (in years) and infant birth year. Controls were selected randomly from children born without Down’s syndrome, as indicated by both the birth record and the BERD database or by birth record only for the years for which BERD was not available. Exact matching on maternal age was used to minimize the potential for residual confounding. For two cases, only four controls could be identified; thus, there were 4,488 controls in total.

All cases and controls were limited to infants with mothers aged 20 years or more (because of the small number of women under age 20 with a parity greater than two) for whom information on parity was available on the birth certificate. Twelve case infants with missing parity information were excluded (1.3 percent of all identified cases); a small proportion from the control pool (2.6 percent) were similarly excluded because of a missing value for parity.

Exposure status was ascertained using parity information from the birth certificate records. Five different levels of parity were examined: zero, one, two, three, and four or more previous livebirths. Additionally, information on potentially confounding factors was obtained from the birth records. Potential confounders that were considered included infant’s race, maternal smoking during pregnancy, and a variety of variables related to the reproductive history of the mother, for both previous pregnancies (prior fetal death or prior induced termination) and the index pregnancy (whether amniocentesis was performed).

Conditional logistic regression was used to assess the risk of Down’s syndrome associated with increasing parity. Odds ratios and their corresponding 95 percent confidence intervals were calculated, using a parity of zero as the reference category. Linear tests of trend (two-sided) were also conducted. Preliminary analyses suggested that the association between parity and Down’s syndrome differed by maternal age; therefore, odds ratios were calculated separately for women under age 35 years and women aged 35 years or more. Likelihood ratio tests were used to determine the significance of the interaction between maternal age and parity (two-sided tests with an alpha level of 0.05). All analyses were carried out using the statistical software package Stata, version 7.0 (Stata Corporation, College Station, Texas).

Because women of higher parity may have a lower likelihood of choosing to terminate a pregnancy in which they are carrying a Down’s syndrome fetus (6, 8, 1113), we conducted a second analysis in which Down’s syndrome infants and controls were excluded if amniocentesis during the index pregnancy was reported on the birth certificate. Women not receiving amniocentesis would probably not know of any genetic abnormalities in the fetus, and thus any association between parity and Down’s syndrome in this group would be less likely to be due to differential likelihood of pregnancy termination. For this analysis, odds ratios and 95 percent confidence intervals were calculated and trend tests and likelihood ratio tests were conducted as above. This subanalysis was based on information from 763 cases and 3,276 controls.


    RESULTS
 TOP
 ABSTRACT
 INTRODUCTION
 MATERIALS AND METHODS
 RESULTS
 DISCUSSION
 REFERENCES
 
The prevalence of Down’s syndrome in Washington State for the entire study period (as measured by both birth certificates and BERD) was 8.1 per 10,000 livebirths. For the years in which BERD was available, the prevalence ranged from a low of 7.5 per 10,000 livebirths in 1990 to a high of 11.2 per 10,000 livebirths in 1989. There was a trend towards a decreasing prevalence of Down’s syndrome over the course of the study. For three of the study years, 1987–1989, a population-based birth defects surveillance program monitored the prevalence of Down’s syndrome in Washington State, reporting a prevalence of 11.3 per 10,000 livebirths for this period (1). Using birth certificates and BERD, we estimated a prevalence of 10.6 per 10,000 livebirths for the same period.

Selected characteristics of cases and controls are shown in table 1. Approximately one third of Down’s syndrome cases were born to mothers over age 35. Even after age-matching, case mothers were more likely to be of higher parity than controls; this relation was particularly strong among older women (aged 35 years or more). Patterns of amniocentesis usage also differed by maternal age. Older control mothers were much more likely to receive amniocentesis than mothers of Down’s syndrome infants. This association was reversed in younger women (under age 35 years), with a slight excess in utilization among case mothers. Case mothers were also slightly less likely to report a previous induced pregnancy termination than control mothers.


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TABLE 1. Selected characteristics of mothers of children with and without Down’s syndrome, Washington State, 1984–1998
 
Older mothers’ use of amniocentesis during the index pregnancy also differed by parity (figure 1). Among control women aged 35 years or more, there was a clear trend towards a decreasing proportion of women receiving amniocentesis with increasing parity. This trend existed despite the fact that higher-parity women were, on average, slightly older than low-parity women (the mean age was 39.8 years for control mothers with a parity of four or greater versus 38.3 years for controls with a parity of zero). No such trend in amniocentesis utilization existed among mothers of Down’s syndrome infants. In mothers under age 35 years, there was no strong pattern of amniocentesis usage based on parity (data not shown).



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FIGURE 1. Use of amniocentesis during pregnancy by women aged 35 years or more among mothers of Down’s syndrome cases and controls, by parity, Washington State, 1984–1998.

 
For all conditional logistic regression models, inclusion of additional covariates did not appreciably change the odds ratios, and thus all odds ratios were adjusted for age only. Odds ratios for the association between Down’s syndrome and parity for all study subjects are shown in table 2. There was an increasing risk of Down’s syndrome with increasing parity for both older and younger women. However, the association between parity and Down’s syndrome was stronger in older women (p for interaction = 0.02 by likelihood ratio test). The risk of Down’s syndrome was nearly four times as great in older women with a parity of four or greater (compared with nulliparous women); in contrast, younger women with a parity of three or more experienced only an approximately 50 percent increase in risk.


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TABLE 2. Age-adjusted odds ratios for Down’s syndrome according to parity and maternal age, Washington State, 1984–1998
 
Restriction of the analysis to women without an indication of amniocentesis on the birth certificate resulted in attenuation of the odds ratios among women aged 35 years or more (table 3). The greatest risk was still observed in older women with a parity of four or more (odds ratio = 2.41, 95 percent confidence interval: 1.41, 4.12), and there remained a trend towards increasing risk with increasing parity. The results for younger women did not change substantially when the analysis was restricted to those without an indication of amniocentesis. Overall, there was a stronger association between parity and Down’s syndrome among older women; however, the interaction between parity and maternal age was no longer statistically significant (p = 0.15 by likelihood ratio test).


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TABLE 3. Age-adjusted odds ratios for Down’s syndrome according to parity and maternal age for women without an indication of amniocentesis on the birth certificate, Washington State, 1984–1998
 
For all of the above conditional logistic regression models, similar trends with parity remained in analyses limited to the years for which BERD was available or in subanalyses that considered only cases identified by particular source(s) (birth certificate only, BERD only, or both). Furthermore, similar patterns were observed if the analysis was restricted to different periods of time within the study years (1984–1986, 1987–1990, 1991–1994, and 1995–1998).


    DISCUSSION
 TOP
 ABSTRACT
 INTRODUCTION
 MATERIALS AND METHODS
 RESULTS
 DISCUSSION
 REFERENCES
 
Higher parity was associated with an increased risk of giving birth to a Down’s syndrome infant, both for women under 35 years of age and for women aged 35 years or more, after results were controlled for maternal age. This association was particularly strong among older women. The exclusion of study subjects with a history of amniocentesis during the index pregnancy blunted the association between parity and Down’s syndrome in older women. Because older women of higher parity in this population were less likely to receive amniocentesis (and therefore less likely to electively terminate a Down’s syndrome pregnancy), part of the observed association seen without the exclusion in place was due to social rather than biologic factors. Higher-parity and case women appear to have somewhat different attitudes towards prenatal diagnosis and pregnancy termination than lower-parity and control women, as suggested by a reduced likelihood of amniocentesis utilization and a decreased number of women reporting prior induced terminations.

There are several limitations to using birth certificate data that must be considered when interpreting the results of this study. First, birth defects are underreported on birth certificates, even after the conversion of open-ended questions to a check-box format on the 1989 revision of the US Standard Certificate of Live Birth (14). In a validation study using Georgia birth certificates from 1989 and 1990, the sensitivity of Down’s syndrome reporting was only 19 percent in comparison with a population-based birth defects registry (15). We attempted to minimize any potential bias of Down’s syndrome underreporting by identifying additional cases of Down’s syndrome using BERD. The addition of BERD data seemed to identify a majority of the cases not reported on the birth certificate, at least for the years for which statewide surveillance data were available. Our estimated prevalence of Down’s syndrome was similar to the prevalence reported by a population-based birth defects registry for the years 1987–1989 (1). Furthermore, the source of case identification did not seem to affect the association with parity, since trends were observed regardless of whether the cases were identified by birth certificate only, by BERD only, or by both methods.

Secondly, we were unable to exclude all individuals who had received prenatal diagnosis for Down’s syndrome from the subgroup analysis. Previous studies have also shown amniocentesis to be underreported on birth certificates. In an evaluation of Tennessee birth certificates from 1989, the sensitivity of amniocentesis reporting ranged from approximately 50 percent (for very low birth weight babies) to 60 percent (for normal weight babies) in comparison with abstraction of subjects’ medical records (16). Additionally, even if reporting of amniocentesis were completely accurate, we failed to exclude anyone who received prenatal diagnosis by chorionic villus sampling, which is not reported on the birth certificate. However, in analysis of 1994–1998 data from the statewide system of regional genetic clinics for Washington State, amniocentesis accounted for more than 95 percent of procedures performed for prenatal diagnosis of Down’s syndrome (D. Lochner Doyle, Washington State Department of Health, personal communication, 2001).

Two different strategies were used to account for the differential use of prenatal diagnosis between low- and high-parity women. First, we examined the risk of Down’s syndrome among multiparous women under the age of 35 years. Prenatal diagnosis is performed infrequently in this group; in our data, amniocentesis was reported in only 2 percent of control mothers under 35 years of age. Therefore, differential elective termination of Down’s syndrome fetuses would be unlikely to result in an appreciable bias in the odds ratios. The second strategy was to perform a subanalysis restricted to those women who had not received amniocentesis during the index pregnancy, according to the birth certificate. This restriction resulted in a decrease in the odds ratios for older women. Had we been able to identify all women receiving prenatal diagnosis, it seems likely that the odds ratios for older women would have been further blunted, becoming even more similar to the odds ratios observed for younger women. In fact, the interaction between parity and age was no longer significant after women reporting amniocentesis were excluded, despite the fact that we were unable to eliminate the impact of the diagnosis and termination of Down’s syndrome pregnancies. Thus, these data seem consistent with a moderate biologic association between parity and Down’s syndrome that does not differ by maternal age and with the possibility that the interaction with age that we observed was due to the differential use of prenatal diagnosis among older women. However, it was impossible to verify this without having complete ascertainment of prenatal diagnosis usage.

Prior studies of parity and Down’s syndrome have not reported associations as strong as those observed in this study. Several studies (25) reported more modest associations with parity, despite the use of 5-year age categories to control for confounding. The use of broader maternal age intervals would tend to bias the resulting measures of association upwards, because of residual confounding. Of the two studies that have controlled for exact maternal age (in years) (6, 8), only one found an association with parity (6), and this was restricted to women aged 35 years or more, leading the authors to conclude that differential usage of prenatal diagnosis between low- and high-parity women was probably responsible for the association.

In attempts to remove the effects of differential pregnancy termination, two previous studies of parity and Down’s syndrome risk have used information from both pregnancies that resulted in livebirths and pregnancies that did not. The first study, by Haddow and Palomaki (7), used results from second-trimester karyotype analysis of 54 cases of Down’s syndrome and 5,282 pregnancies with a normal karyotype in a cohort of women aged 35 years or more in Minnesota. In the second, conducted in Australia, Chan et al. (8) had access to a birth defects registry that included karyotype results from both liveborn and electively aborted fetuses. Their analysis was based on 284 cases of Down’s syndrome and almost 200,000 livebirths. Interestingly, neither of these studies reported an association between parity and Down’s syndrome after controlling for maternal age, though only one (8) calculated an age-adjusted relative risk (for each additional livebirth, the relative risk was 0.96 (95 percent confidence interval: 0.78, 1.20)). Another study that examined gravidity of four or more as a risk factor for Down’s syndrome (17) noted a reduction in odds ratios (from 1.41 to 1.15) when spontaneous fetal losses and terminations were included (as compared with livebirths only).

One potential drawback of using information from pregnancies that do not result in livebirth in this type of analysis is that a large majority of Down’s syndrome fetuses, usually an estimated 75 percent or more (18, 19), are spontaneously aborted. At the time of elective termination of a Down’s syndrome fetus, it is impossible to know whether the pregnancy would have resulted in a livebirth. If the association between parity and Down’s syndrome were due to a reduced likelihood of pregnancy loss by women of higher parity, as has been proposed (10, 20), the inclusion of all Down’s syndrome pregnancies that did not result in livebirth would eliminate any association, even if one were truly present. Women of proven ability to carry a pregnancy to completion, as measured by number of prior livebirths, may be more likely to do so even in the presence of fetal chromosomal abnormalities. If this were the case, it could explain part of the seemingly contradictory results between our analyses and those of Haddow and Palomaki (7) and Chan et al. (8), since the latter studies included some cases of Down’s syndrome in which spontaneous abortion would have occurred had the pregnancies been allowed to continue. If spontaneous loss of a Down’s syndrome fetus were more likely to occur among women of lower parity, the inclusion of case fetuses that were destined to be spontaneously aborted would bias any true association towards the null.

Still, the inclusion of these cases is unlikely to account for all of the differences between our analyses and those of Haddow and Palomaki (7) and Chan et al. (8). Down’s syndrome pregnancy loss is believed to be more common earlier in gestation. After the time of amniocentesis (usually 15–18 weeks’ gestation), loss rates have been estimated to be 10–25 percent (2123). Thus, at most, one fourth of the case fetuses in the Haddow and Palomaki study and one tenth of the case fetuses in the Chan et al. study (one fourth of the 113 terminated cases included) would have been spontaneously aborted. This is an insufficient number of cases to fully account for the differences from our study. Nevertheless, if spontaneous Down’s syndrome fetal loss differs by parity, our method of accounting for differential termination practices by restricting the data to women not receiving prenatal diagnosis is not susceptible to the particular bias introduced by the inclusion of terminated cases. In practice, however, we were hindered in carrying out this approach because of underreporting of prenatal diagnosis on the birth certificate.

It is unclear why our findings differ from those of Haddow and Palomaki (7) and Chan et al. (8). We cannot rule out the possibility of uncontrolled confounding or other bias leading to a spurious association. However, the magnitude of the association observed in this study, even among younger women, is strong enough that additional studies in populations in which prenatal diagnosis may be more reliably ascertained are warranted.


    ACKNOWLEDGMENTS
 
Support for this work was provided in part by grant 5T32CA09168 from the National Institutes of Health.

The authors thank Drs. Christine Velicer and Noel Weiss for their helpful comments on an earlier version of the manuscript and William O’Brien for his assistance with the birth record and BERD databases.


    NOTES
 
Correspondence to Dr. V. Paul Doria-Rose, Division of Public Health Sciences, Fred Hutchinson Cancer Research Center, 1100 Fairview Avenue North, MP-381, P.O. Box 19024, Seattle, WA 98109-1024 (e-mail: pdoriaro{at}fhcrc.org). Back


    REFERENCES
 TOP
 ABSTRACT
 INTRODUCTION
 MATERIALS AND METHODS
 RESULTS
 DISCUSSION
 REFERENCES
 

  1. Down syndrome prevalence at birth—United States, 1983–1990. MMWR Morb Mortal Wkly Rep 1994;43:617–22.[Medline]
  2. Eidelman AI, Kamar R, Schimmel MS, et al. The grandmultipara: is she still a risk? Am J Obstet Gynecol 1988;158:389–92.[ISI][Medline]
  3. Schimmel MS, Eidelman AI, Zadka P, et al. Increased parity and risk of trisomy 21: review of 37,110 live births. BMJ 1997;314:720–1.[Free Full Text]
  4. Castilla EE, Paz JE. Parity and Down’s syndrome. (Letter). Lancet 1994;344:1645–6.
  5. Kallen B, Masback A. Down syndrome: seasonality and parity effects. Hereditas 1988;109:21–7.[ISI][Medline]
  6. Kallen K. Parity and Down syndrome. Am J Med Genet 1997;70:196–201.[CrossRef][ISI][Medline]
  7. Haddow JE, Palomaki GE. Multiparity and Down’s syndrome. (Letter). Lancet 1994;344:956.[ISI][Medline]
  8. Chan A, McCaul KA, Keane RJ, et al. Effect of parity, gravidity, previous miscarriage, and age on risk of Down’s syndrome: population based study. BMJ 1998;317:923–4.[Free Full Text]
  9. Rogers MS. Racial variations in the incidence of trisomy 21. Br J Obstet Gynecol 1986;93:597–9.[ISI][Medline]
  10. Lilford RJ. Commentary: Down’s syndrome and parity. BMJ 1997;314:721.[Free Full Text]
  11. Halliday J, Lumley J, Watson L. Comparison of women who do and do not have amniocentesis or chorionic villus sampling. Lancet 1995;345:704–9.[ISI][Medline]
  12. Lesser Y, Rabinowitz J. Elective amniocentesis in low-risk pregnancies: decision making in the era of information and uncertainty. Am J Public Health 2001;91:639–41.[Abstract]
  13. Roghmann KJ, Doherty R, Robinson JL, et al. The selective utilization of prenatal genetic diagnosis: experiences of a regional program in Upstate New York during the 1970s. Med Care 1983;21:1111–25.[ISI][Medline]
  14. Freedman MA, Gay GA, Brockert JE, et al. The 1989 revisions of the US Standard Certificates of Live Birth and Death and the US Standard Report of Fetal Death. Am J Public Health 1988;78:168–72.[Abstract]
  15. Watkins ML, Edmonds L, McClearn A, et al. The surveillance of birth defects: the usefulness of the revised US standard birth certificate. Am J Public Health 1996;86:731–4.[Abstract]
  16. Piper JM, Mitchel EF Jr, Snowden M, et al. Validation of 1989 Tennessee birth certificates using maternal and newborn hospital records. Am J Epidemiol 1993;137:758–68.[Abstract]
  17. Torfs CP, Christianson RE. Effect of maternal smoking and coffee consumption on the risk of having a recognized Down syndrome pregnancy. Am J Epidemiol 2000;152:1185–91.[Abstract/Free Full Text]
  18. Hook EB. Epidemiology of Down syndrome. In: Pueschel SM, Rynders JE, eds. Down syndrome: advances in biomedicine and the behavioral sciences. Cambridge, MA: Ware Press, 1982:11–88.
  19. Boue J, Deluchat CC, Nicolas H, et al. Prenatal losses of trisomy 21. In: Burgio GR, Fraccaro M, Tiepolo L, et al, eds. Trisomy 21: an international symposium. Berlin, Germany: Springer-Verlag, 1981:183–93.
  20. Pharoah PO. Increased parity and risk of trisomy 21: study measured prevalence of Down’s syndrome at birth, not incidence. (Letter). BMJ 1997;314:1760–1.[Free Full Text]
  21. Bray IC, Wright DE. Estimating the spontaneous loss of Down syndrome fetuses between the times of chorionic villus sampling, amniocentesis and livebirth. Prenat Diagn 1998;18:1045–54.[CrossRef][ISI][Medline]
  22. Morris JK, Wald NJ, Watt HC. Fetal loss in Down syndrome pregnancies. Prenat Diagn 1999;19:142–5.[CrossRef][ISI][Medline]
  23. Cuckle H. Down syndrome fetal loss rate in early pregnancy. (Letter). Prenat Diagn 1999;19:1177–9.

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Doria-Rose and Edwards Respond to "Parity and Down’s Syndrome"
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