Reexamination of paternal age effect in Down's syndrome

Summary The recent discovery that the extra chromosome in about 30% of cases of 47, trisomy 21 is of paternal origin has revived interest in the possibility of paternal age as a risk factor for a Down syndrome birth, independent of maternal age. Parental age distribution for 611 Down's syndrome 47,+21 cases was studied. The mean paternal age was 0.16 year greater than in the entire population of live births after controlling for maternal age. There was no evidence for a significant paternal age effect at the 0.05 level. For 242 of these Down's syndrome cases, control subjects were selected by rigidly matching in a systematic manner. Paternal age was the variable studied, with maternal age and time and place of birth controlled. There was no statistically significant association between paternal age and Down's syndrome. After adjustment for maternal age, these two studies were not consistent with an increase of paternal age in Down's syndrome.     

The meiotic stage of nondisjunction in trisomy 21: Determination by using DNA polymorphisms

We have studied DNA polymorphisms at loci in the pericentromeric region on the long arm of chromosome 21 in 200 families with trisomy 21, in order to determine the meiotic origin of nondisjunction. Maintenance of heterozygosity for parental markers in the individual with trisomy 21 was interpreted as resulting from a meiosis I error, while reduction to homozygosity was attributed to a meiosis II error. Nondisjunction was paternal in 9 cases and was maternal in 188 cases, as reported earlier. Among the 188 maternal cases, nondisjunction occurred in meiosis I in 128 cases and in meiosis II in 38 cases; in 22 cases the DNA markers used were uninformative. Therefore meiosis I was responsible for 77.1% and meiosis II for 22.9% of maternal nondisjunction. Among the 9 paternal nondisjunction cases the error occurred in meiosis I in 2 cases (22.2%) and in meiosis II in 7 (77.8%) cases. Since there was no significant difference in the distribution of maternal ages between maternal I error versus maternal II error, it is unlikely that an error at a particular of maternal ages between maternal I error versus maternal II error, it is unlikely that an error at a particular meiotic stage contributes significantly to the increasing incidence of Down syndrome with advancing maternal age. Although the DNA polymorphisms used were at loci which map close to the centromere, it is likely that rare errors in meiotic-origin assignments may have occurred because of a small number of crossovers between the markers and the centromere.(ABSTRACT TRUNCATED AT 250 WORDS)     

Down syndrome rates and relaxed selection at older maternal ages.

Preferential survival in older mothers of fetuses with Down syndrome has been proposed as contributing to the maternal-age effect of this condition. If correct, this provocative hypothesis, which may be termed "relaxed selection," has major implications for approaches to prevention of Down syndrome live births in older women. Several predictions of this hypothesis are examined here by comparisons of parental ages among various populations. These revealed that: (1) mean maternal age of Down syndrome live births is slightly lower than that of Down syndrome spontaneous fetal deaths; (2) mean maternal age of those with mutant D/21 translocation Down syndrome is about the same as that of controls; (3) the ages of Down syndrome mothers who have Down syndrome live births is slightly lower than ages of Down syndrome mothers who have unaffected live births; and (4) in recent data on 47, +21 cases in which the extra chromosome 21 is of paternal origin, the mean maternal ages are 4-5 years lower than the maternal ages of cases of maternal origin (in contrast to earlier reports). All of these observations are contrary to the hypothesis that relaxed selection contributes significantly to the maternal-age association of Down syndrome. If there is any effect of relaxed selection, it is likely to be very weak and/or act primarily upon abortions that occur before recognition of pregnancy.     

Parental origin of the extra chromosome in prenatally diagnosed fetal trisomy 21

Trisomy 21 (Down syndrome) is one of the most common chromosomal abnormalities. Of cases of free trisomy 21 causing Down syndrome, about 95% result from nondisjunction during meiosis, and about 5% are due to mitotic errors in somatic cells. Previous studies using DNA polymorphisms of chromosome 21 showed that paternal origin of trisomy 21 occurred in only 6.7% of cases. However, these studies were conducted in liveborn trisomy 21-affected infants, and the possible impact of fetal death was not taken into account. Using nine distinct DNA polymorphisms, we tested 110 families with a prenatally diagnosed trisomy 21 fetus. Of the 102 informative cases, parental origin was maternal in 91 cases (89.2%) and paternal in 11 (10.8%). This percentage differs significantly from the 7.0% observed in previous studies (P<0.001). In order to test the influence of genomic parental imprinting, we determined the origin of the extra chromosome 21 in relation to different factors: advanced maternal age, maternal serum human chorionic gonadotropin (hormone of placental origin), severity of the disease, gestational age at diagnosis and fetal gender. We found that the increased frequency of paternal origin of nondisjunction in trisomy 21-affected fetuses cannot obviously be explained by factors leading to selective loss of paternal origin fetuses.     

Parental origin and meiotic stage of non-disjunction in 139 cases of trisomy 21

The parental origin and the meiotic stage of non-disjunction have been determined in 139 Down syndrome patients with regular trisomy 21 and in their parents through the analysis of DNA polymorphism. The meiotic error is maternal in 91.60% cases and paternal in 8.39% of cases. Of the maternal cases, 72.41% were due to meiosis I errors (MMI) and 27.58% were due to meiosis II errors (MMII). Of the paternal cases, 45.45% were due to meiosis I (PMI) and 54.54% were due to meiosis II (PMII). The mean maternal ages were 31.6 +/- 5.3 (+/- SD) years in errors from MMI, 32.3 +/- 6.4 years in errors from MMII, 31.4 +/- 4.6 years in errors from PMI and 29.5 +/- 2.7 years in errors from PMII. No significant statistical differences were observed between maternal and paternal errors, further supporting the presence of a constant chromosome 21 non-disjunction error type.     

Analysis of DNA polymorphisms suggests that most de novo dup(21q) chromosomes in patients with Down syndrome are isochromosomes and not translocations.

Down syndrome is rarely due to a de novo duplication of chromosome 21 [dup(21q)]. To investigate the origin of the dup(21q) and the nature of this chromosome, we used DNA polymorphisms in 10 families with Down syndrome due to de novo dup(21q). The origin of the extra chromosome 21q was maternal in six cases and paternal in four cases. Furthermore, the majority (eight of 10) of dup(21q) chromosomes were isochromosomes i(21q) (four were paternal in origin, and four were maternal in origin); however, in two of 10 families the dup(21q) chromosome appeared to be the result of a Robertsonian translocation t(21q;21q) (maternal in origin in both cases).     

Issues in analysis of data on paternal age and 47,+21: implications for genetic counseling for Down syndrome

Summary Data and analyses on paternal age and 47,+21 are reviewed. It is concluded that there are few, if any, grounds to justify the inference of a paternal age effect independent of maternal age for those paternal age-maternal age combinations on which there are prenatal diagnostic data. It is suggested that genetic counseling as to increased (or decreased) risk of Down syndrome associated with various paternal ages is not justified at present.     

Paternal age and Down's syndrome genotypes diagnosed prenatally: No association in New York State data

Summary An investigation of a paternal age effect independent of maternal age was undertaken for 98 cases of Down's syndrome genotypes diagnosed prenatally compared to 10,329 fetuses with normal genotype diagnosed prenatally in data reported to the New York State Chromosome Registry. The mean of the difference (delta) in paternal age of cases compared to those with normal genotypes after controlling for maternal age, was slightly negative,-0.27 with a 95% confidence interval of-1.59 to +1.06. A regression analysis was also done in which the data were first fit to an equation of the type lny=(bx+c) and then to the equation ln y=(bx+dz+c) where y = rate of Down's syndrome, x = maternal age, z = paternal age, and b, d, and c are parameters. This also revealed no evidence for a paternal age effect. The value of d (the paternal age coefficient) was in fact slightly negative,-0.0058, with an asymptotic 95% confidence interval of-0.0379 to +0.0263. Lastly, multiple applications of the Mantel-Haenszel test considering various boundaries in paternal age also revealed no statistically significant evidence for a paternal age effect independent of maternal age. These results are at variance with claims of others elsewhere of a very strong paternal age effect detected in studies at prenatal diagnoses. Five different hypotheses are suggested which may account for discrepancies among studies to date in findings on paternal age effects for Down's syndrome: (i) there are temporal, geographic, or ethnic variations in paternal age effects, (ii) there is no paternal age effect and statistical fluctuation accounts for all trends to date; (iii) methologic artifacts have obscured a paternal age effect in some studies which did not find a positive outcome; (iv) methodologic artifacts are responsible for the positive results in some studies to date; (v) there is a rather weak paternal age effect independent of maternal age in most if not all populations, but because of statistical fluctuation the results are significant only in some data sets. The results of all data sets to date which we have been able to analyze by one year intervals are consistent with a mean delta of +0.04 to +0.48 and in the value of d (the paternal age coefficient) of +0.006 to +0.017, and it appears the fifth hypothesis cannot be excluded. Projections based on this assumption are presented.     

Translocation Down syndrome in Ohio 1970-1981: epidemiologic and cytogenetic factors and mutation rate estimates.

Sixteen hundred eighty-eight Down syndrome live births, including 65 (5.2%) translocations, were ascertained in Ohio between 1970 and 1981. Translocations of known origin were 24.4% maternal, 2.2% paternal, and 73.3% de novo. Translocation subtypes were 14/21 (45.7%), 15/21 (2.9%), 21/21 (40.0%), 21/22 (2.9%), and other (8.5%). Among 14/21 translocations, 33.3% were maternal in origin and 66.7% were de novo, while 100% of 21/21 translocations were de novo. No differences were found when the maternal- and paternal-age distributions of all translocations or various translocation subsets were compared with the live-birth control distributions. However, mean maternal and paternal ages of de novo translocations were significantly lower than that of the live-birth controls. Ohio data showed the average maternal age of de novo D/21 cases to be significantly lower than the control. Ages of both parents of de novo G/21 cases and paternal age of D/21 cases were not different from the control. De novo translocation mutation rate estimates were 0.8 X 10(-5) for 14/21, 1.2 X 10(-5) for 21/21, and 2.2 X 10(-5) overall. Ohio estimates (3.2 X 10(-5) for 1970-1972 and 1.4 X 10(-5) for 1973-1975) did not reflect the increase in mutation rate previously found in New York during 1973-1977.     

Grandmaternal age in children with Down syndrome in St. Petersburg

Advanced maternal age is a well-established factor of DS occurrence. However the majority of DS cases are born to young couples. Some studies suggested that the risk for Down syndrome may be related to an aging grandmother. We obtained data on grandmaternal ages in 243 families of DS and 330 families of healthy children born in 1990-1999. The data were analyzed according to two categories of maternal ages, <30 yr and > or =30 yr. We did not find systematic differences in grandparental age distribution between the studied groups. Specifically, in 102 young couples with DS, medians for both maternal and paternal grandmother's age appeared to be equal (26 yr). Similar figures were observed in 284 young controls (27 yr). There was no difference in age distribution between 141 older couples with DS and 104 control couples. Therefore we failed to support the suggestion that advanced age of the DS grandmother is responsible for meiotic disturbance in her daughter. Neither the hypothesis suggesting a significant contribution of parentally transmitted trisomy 21 to DS population rate has been confirmed.     

Maternal cigarette smoking, Down syndrome in live births, and infant race.

Previous studies have suggested that maternal smoking is negatively associated with a Down syndrome live birth. We analyzed the data of the U.S. Perinatal Collaborative Study in a search for racial variation in Down syndrome risk factors. There were 22 cases in 25,346 live births to smoking mothers (4/10,780 blacks, 18/13,320 whites, and 0/1,246 other races) and 42/29,130 live births to nonsmoking mothers (24/14,665 blacks, 14/11,694 whites, and 4/2,771 others). The crude overall rates per 1,000 live births were 0.4 in black smokers and 1.6 in black nonsmokers but 1.4 in white smokers and 1.2 in white non-smokers. Adjusted for maternal age, the summary relative risk for a Down syndrome live birth to a smoking mother was 0.2 in blacks (95% interval 0.1-0.7) but 1.2 in whites (95% interval 0.6-2.5). Stratification on variables associated with socioeconomic status or gestational age at time of entry into the study did not alter the racial difference. A comparison of smokers with those who never smoked revealed essentially the same trends. Among all nonsmokers the ratio of the maternal age-adjusted risks for a Down syndrome live birth in whites compared with blacks was 0.7 (95% interval 0.3-1.3), and among all smokers this ratio was 3.6 (95% interval 1.3-9.9). If the results are not attributable to statistical fluctuation or undetected confounding, then differences in the probability of intrauterine survival of the Down syndrome fetus would appear to be one plausible explanation for the difference.     

Epidemiology of Down syndrome in South Australia, 1960-89.

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 less than .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.     

Prenatal diagnosis, pregnancy terminations and prevalence of Down syndrome in Atlanta

BACKGROUND: The impact of prenatal diagnosis on the live birth prevalence of Down syndrome (trisomy 21) has been described. This study examines the prevalence of Down syndrome before (1990-1993) and after inclusion of prenatally diagnosed cases (1994-1999) in a population-based registry of birth defects in metropolitan Atlanta. METHODS: We identified infants and spontaneous fetal deaths with Down syndrome (n = 387), and pregnancies electively terminated after a prenatal diagnosis of Down syndrome (n = 139) from 1990 to 1999 among residents of metropolitan Atlanta from a population-based registry of birth defects, the Metropolitan Atlanta Congenital Defects Program (MACDP). Only diagnoses of full trisomy 21 were included. Denominator information on live births was derived from State of Georgia birth certificate data. We compared the prevalence of Down syndrome by calendar period (1990-1993, 1994-1999), maternal age (<35 years, 35+ years), and race/ethnicity (White, Black, other), using chi-square and Fisher's exact tests. RESULTS: During the period when case ascertainment was based only on hospitals (1990-1993), the prevalence of Down syndrome was 8.4 per 10,000 live births when pregnancy terminations were excluded and 8.8 per 10,000 when terminations were included. When case ascertainment also included perinatal offices (1994-1999), the prevalence of Down syndrome was 10.1 per 10,000 when terminations were excluded and 15.3 when terminations were included. During 1990-1993, the prevalence of Down syndrome was 24.7 per 10,000 among offspring to women 35+ years of age compared to 6.8 per 10,000 among offspring to women <35 years of age (rate ratio [RR] = 3.65, 95% confidence interval [CI] = 2.53-5.28). During 1994-1999, the prevalence of Down syndrome was 55.3 per 10,000 among offspring to women 35+ years compared to 8.5 per 10,000 among offspring to women <35 years (RR = 6.55, 95% CI = 5.36-7.99). There was no statistically significant variation in the prevalence of Down syndrome by race/ethnicity within maternal age and period of birth strata. During 1994-1999, the proportion of cases that were electively terminated was greater for women 35+ years compared to women <35 years (RR = 5.10, 95% CI = 3.14-8.28), and lower for Blacks compared to Whites among women 35+ years of age (RR = 0.33, 95% CI = 0.16-0.66). CONCLUSIONS: In recent years, perinatal offices have become an important source of cases of Down syndrome for MACDP, contributing at least 34% of cases among pregnancies in women 35+ years of age. Variation in the prevalence of Down syndrome by race/ethnicity, before or after inclusion of cases ascertained from perinatal offices, was not statistically significant. Among Down syndrome pregnancies in mothers 35+ years we found a lower proportion of elective termination among Black women compared to White women. We suggest that future reports on the prevalence of Down syndrome by race/ethnicity take into account possible variations in the frequency of prenatal diagnosis or elective termination by race/ethnicity.     

An analysis for paternal-age effect in Ohio's Down syndrome births, 1970-1980.

The purpose of this study was to analyze Down syndrome (DS) births during 1970-1980 in the State of Ohio for a paternal-age effect independent of maternal age. Birth certificates and chromosome analysis records were used to ascertain 1,244 white DS births, which by capture-recapture methodology were estimated to comprise two-thirds of all white DS births in Ohio for this period. The control data consisted of 1,667,210 white live births in Ohio during the same period. One method of statistical analysis was a case-control comparison, which for each single-year maternal age compares the mean paternal age for controls with each observed DS paternal age. No statistically significant paternal-age effect was found in nine of the 11 years. For two of the years, and for all years combined, the DS fathers were significantly younger than the fathers of controls. When the data were subdivided according to ascertainment, one subpopulation--those DS individuals obtained from birth certificates alone--also showed a statistically significant negative paternal-age effect. The Mantel-Haenszel test was also applied to these data. Assuming no paternal-age effect, a lower rate of DS births than expected was found at paternal ages greater than or equal to 40, but not at greater than or equal to 45, greater than or equal to 50, or greater than or equal to 55. These same methods were used to test for a maternal-age effect. In each of the 11 years and over all 11 years combined, a strong and statistically significant positive maternal-age effect was detected.     

Epidemiology of Down syndrome (Trisomy 21), Hawaii, 1986-97

BACKGROUND: The live birth prevalence of Down syndrome is approximately 10 per 10,000 live births in the United States. Down syndrome prevalence has been reported to change over time and to vary by selected demographic factors. METHODS: Data from a population-based birth defects registry in Hawaii involving 363 Down syndrome cases delivered during 1986-97 were used to calculate overall prevalence and to investigate secular trends and differences by selected demographic factors. RESULTS: The total (live birth, fetal death, and elective termination) prevalence was 14.74 per 10,000 live births and fetal deaths. The unadjusted live birth prevalence was 8.67 per 10,000 live births. The adjusted live birth (live births and proportion of elective terminations expected to have resulted in live births) prevalence was 12.59 per 10,000 live births. No significant secular trends were observed for either total prevalence (P = 0.688) or adjusted live birth prevalence (P = 0.604). The total Down syndrome prevalence per 10,000 live births was highest for Far East Asians (22.01), followed by whites (17.06), Filipinos (15.94), and Pacific Islanders (9.21). Prevalence per 10,000 births was higher in metropolitan Honolulu (18.57) than in the rest of Hawaii (14.15). After adjusting for maternal age, however, the differences within the demographic groups were not statistically significant. CONCLUSIONS: The live birth prevalence of Down syndrome in Hawaii during 1986-97 was lower than reported in the literature. Prevalence did not change significantly over time. Any differences in prevalence by maternal race/ethnicity and place of residence appeared to result from differences in maternal age distribution.     

Paternal occupational exposures and the risk of Down syndrome.

An exploratory case-control study of paternal occupation as a risk factor for Down syndrome was conducted. With the use of the British Columbia Health Surveillance Registry, 1,008 cases of live-born Down syndrome were identified for the period 1952-73. Two controls were matched to each case by using the birth files of British Columbia. Paternal occupation was obtained from the birth notice. Elevated maternal age-adjusted relative risks of Down syndrome were found for fathers employed as janitors (odds ratio [OR] = 3.26; 95% confidence interval [C.I.] = 1.02-10.44); mechanics (OR = 3.27; C.I. = 1.57-6.80); farm managers/workers (OR = 2.03; C.I. = 1.25-3.03); material-moving equipment operators (OR = 1.88; C.I. = 0.93-3.82); food processors (OR = 1.79; C.I. = 0.96-3.31); sheet-metal workers, iron workers, and other metalworkers (OR = 1.57; C.I. = 0.92-2.69); and sawmill workers (OR = 1.43; C.I. = 0.90-2.66). This large study provides new leads for further evaluation of the role of paternal exposures in the etiology of Down syndrome.     

Origin of trisomy 21 in Down syndrome cases from a Spanish population registry

We have carried out a population-based study on the origin of the extra chromosome 21 in 38 families with Down syndrome (DS) offspring in El Vallès (Spain). From 1991 to 1994, a higher prevalence of DS (22.7/10000 live births, stillbirths and induced abortions) was found compared to the majority of EUROCAT registries. The distribution of trisomy 21 by origin was 88% maternal (90.6% meiosis I, 6.2% meiosis II, 3.1% maternal mosaicism), 5.6% paternal (50% meiosis I, 50% meiosis II) and 5.6% mitotic. The percentage of parental mosaicism was 2.7%. These percentages are similar to those previously reported. Recombination study revealed a maternal meiosis I genetic map of 32.68 cM (approximately one-half the length of the normal female map). Mean maternal age among non-recombinant cases involving MI errors was significantly lower (31.1 years) than among those cases showing one observable crossover (36.1 years) (P<0.05); this could support the hypothesis that 'achiasmate' chromosomes may be subject to aberrant segregation regardless of maternal age.     

Paternal age and trisomy among spontaneous abortions

Summary The relationship of paternal age to specific types of trisomy and to chromosomally normal loss was investigated in data drawn from a case-control study of spontaneous abortions. Differences in paternal age between karyotype groups and controls delivering after 28 weeks gestation were tested using an urn model analysis which adjusted, by regression, for maternal age and, by stratification, for the effects of design variables (payment status, phase of study) and demographic factors (language, ethnicity). The magnitude of paternal age differences was estimated using least squares regression analysis. For chromosomally normal cases there was no association with paternal age. Among the fourteen trisomy categories examined, four (7, 9, 18, 21) showed increased paternal age ( 1 year above expectation), three (13, 20, 22) showed decreased paternal age and the rest, including the most common, trisomy 16, showed negligible differences. Only the association with trisomy 22 was statistically significant (P = 0.012), with a predicted reduction in paternal age of 2.1 years (95% CI -4.9, -0.5 years). This association did not vary with maternal age, payment status, phase of study, language or ethnicity. Because previous observations are extensive, the relation of paternal age to trisomy 21 was examined further. The overall association was not significant ( = 0.8 years; 95% CI -0.8, 2.4 years). Moreover, there was evidence that the magnitude and direction of paternal age associations vary significantly within the sample, although not between subgroups defined on the basis of payment, phase of study, language or ethnicity. With respect to maternal age, the trend is towards a greater paternal age difference for trisomy 21 losses in younger women (P = 0.058). Given the number of tests performed, the finding for trisomy 22 and reduced paternal age could be due to chance. Among trisomy types, the direction of paternal age associations was not consistent for chromosomes grouped according to characteristics that might relate to the probability of nondisjunction, such as size, arm ratio, or nucleolar organizer region content, or to the potential viability of the trisomy. Thus, neither on statistical nor biological grounds do the data provide compelling evidence of paternal age effects on the trisomies found among spontaneous abortions, or on chromosomally normal losses.     

Atrioventricular septal defects: possible etiologic differences between complete and partial defects

BACKGROUND: Recent advances in clinical, pathological, and genetic aspects of atrioventricular septal defects (AVSD) have set the stage for epidemiologic investigations into possible risk factors. Previous analyses of the total case group of AVSD included complete and partial subtypes without analysis of the subsets. METHODS: To address the question of possible morphogenetic heterogeneity of AVSD, the Baltimore-Washington Infant Study data on live-born cases and controls (1981-1989) was reanalyzed for potential environmental and genetic risk-factor associations in complete AVSD (n = 213), with separate comparisons to the atrial (n = 75) and the ventricular (n = 32) forms of partial AVSD. RESULTS: Complete and ventricular forms of AVSD had a similar proportion of isolated cases (12.2% and 15.6%, respectively, without associated extracardiac anomalies) and high rates of Down syndrome, whereas the atrial form of partial AVSD included 55% isolated cases. Trisomy 18 occurred in 22% of infants with the ventricular form, compared with <2% in the other AVSD groups. Analysis of potential risk factors revealed further distinctions. Complete AVSD as an isolated cardiac defect was strongly associated with maternal diabetes (odds ratio [OR] = 20.6; 95% confidence interval [CI] =5.6-76.4) and also with antitussive use (OR = 8.8; CI = 1.2-48.2); there were no strong associations other than maternal age among Down syndrome infants with this type of heart defect. Isolated cases with the atrial type of partial AVSD were associated with a family history of heart defects (OR = 6.2; CI = 1.4-24.4) and with paternal occupational exposures to ionizing radiation (OR = 5.1; CI = 1.4-27.4), but no risk factors were associated with Down syndrome. There were no significant associations of any risk factors in the numerically small subsets of isolated and Down syndrome cases with the ventricular form of partial AVSD. CONCLUSIONS: These results indicate a similar risk profile of complete AVSD and the ventricular type of partial AVSD, with a possible subset of the latter due to trisomy 18. Maternal diabetes constituted a potentially preventable risk factor for the most severe, complete form of AVSD.     

Extra structurally abnormal chromosomes (ESAC) detected at amniocentesis: frequency in approximately 75,000 prenatal cytogenetic diagnoses and associations with maternal and paternal age.

We analyzed rates of extra structurally abnormal chromosomes (ESAC) detected in prenatal cytogenetic diagnoses of amniotic fluid reported to the New York Chromosome Registry. These karyotypes include both extra unidentified structurally abnormal chromosomes (EUSAC)--often denoted as "markers"--and extra identified structurally abnormal chromosomes (EISAC). The rate of all EUSAC was 0.64/1,000 (0.32-0.40/1,000 mutant and 0.23-0.32 inherited), and that of all EISAC was 0.11/1,000 (0.07/1,000 mutant and 0.04/1,000 inherited). The rate of all ESAC was approximately 0.8/1,000-0.4-0.5/1,000 mutant and 0.3-0.4/1,000 inherited. Mean +/- SD maternal age of mutant cases was 37.5 +/- 2.9, significantly greater than the value of 35.8 years in controls. A regression analysis indicated a rate of change of the log of the rate of about +0.20 with each year of maternal age between 30 and 45 years. When paternal age was introduced, the maternal age coefficient increased to about +0.25--close to that seen for 47, +21--but the paternal age coefficient was -0.06. After being matched for maternal age and year of diagnosis, the case-control difference in paternal age for 24 mutant cases was -2.4 with a 95% confidence interval of -4.6 to -0.1 years. In a regression analysis of the effects of both parental ages on the (log) rate, the maternal age coefficient was +0.25 and the paternal age coefficient was -0.06. These results are consistent with a (weak) negative paternal age effect in the face of a strong maternal age effect. Since ESAC include a heterogeneous group of abnormalities, the maternal age and paternal age trends, if not the result of statistical fluctuation or undetected biases, may involve different types of events. Data in the literature suggest that chromosomes with de novo duplicated inversions of 15p have a strong maternal age effect (but little paternal age effect). Such chromosomes, however, do not account for the active maternal age trends seen in the data analyzed here. Inherited ESAC exhibited no such trends.