1 Section for Medical Statistics, Department of Public Health and Primary Health Care, University of Bergen, Bergen, Norway.
2 Medical Birth Registry of Norway, University of Bergen, Bergen, Norway.
![]() |
ABSTRACT |
---|
![]() ![]() ![]() ![]() ![]() ![]() ![]() |
---|
abnormalities; birth weight; family; gestational age; growth
Abbreviations: CI, confidence interval
![]() |
INTRODUCTION |
---|
![]() ![]() ![]() ![]() ![]() ![]() ![]() |
---|
It is acknowledged that there is a strong relation between the birth weights of siblings and that growth restriction and preterm delivery recur in sibships (38
). An infant's birth weight is an important predictor for perinatal mortality and morbidity, with an inverse relation between birth weight and perinatal mortality rates (9
12
). However, dependencies within sibships have a modifying effect on this relation, so that the risk of perinatal death for a second-born infant varies with its birth weight relative to that of the older sibling (6
, 13
, 14
).
In a previous study we found that, in families where one infant died in the perinatal period, surviving siblings had significantly lower mean birth weights than did infants of the same birth order in families where all infants survived the perinatal period, except when the cause of death was a congenital anomaly (15). In this latter situation the mean birth weight of surviving siblings was not different from that of corresponding infants in families without perinatal losses.
Our present aim was to study the birth weights of nonmalformed infants in families where one or more of the infants had a congenital malformation and to compare these with the birth weights of infants in families where none had a registered malformation.
![]() |
MATERIALS AND METHODS |
---|
![]() ![]() ![]() ![]() ![]() ![]() ![]() |
---|
The study was based on 551,478 mothers with at least two singleton infants and 209,423 mothers with at least three singletons, with the first infant born in 1967 or later. The mother-sibling units were analyzed as two family sets, so that for mothers with two or more infants the first two were analyzed, and for mothers with three or more infants the first three were analyzed. The family sets were thus not mutually exclusive, but by this design the fallacy of "fixed sibship size" was avoided (16, 17
). Within each set, the families were grouped according to whether and in which birth order an infant was registered as having a congenital malformation (table 1). Families where none of the infants had a registered birth defect were used as reference (control infants).
|
We evaluated confounding by mother's age, mother's educational level, marital status, maternal diabetes, time period, and interpregnancy interval, with the categories as described in tables 2 and 4. Educational level and marital status were used as proxy variables for socioeconomic status (19, 20
).
|
|
In this study categories of birth defects were defined on the basis of the three-digit codes of the International Classification of Diseases, Eighth Revision, with minor modifications, as described previously (2123
). All registered birth defects except congenital hip dislocation were used. For most infants, only a single defect was reported, and it was assumed to be an isolated defect. Multiple defects were combined in a separate category except that, when anencephalus was present with spina bifida, only anencephalus was counted, and when spina bifida was present with hydrocephalus, only spina bifida was counted. In the present paper, we combine anencephalus and spina bifida as neural tube defects. Isolated cleft palate was separated from the three-digit code for cleft lip and palate as a distinct category, and likewise Down's syndrome was separated from other recognized syndromes. Altogether, there were 25 categories of birth defects, one of which contained multiple defects. The most frequent organ-specific malformations were analyzed separately. However, for the main analyses we pooled all birth defects into one group.
In the text we use the terms "birth defect," "(congenital) malformation," and "(congenital) anomaly" interchangeably.
Statistics
We used t tests to compare mean values for infants of the same birth order in the different (independent) family groups. Standard analyses of variance were used to adjust the differences in mean birth weight for gestational age and to evaluate confounding. In these models the independent variables were treated as categorical fixed factors (tables 2 and 4), and each birth order was analyzed separately. To include all siblings in the same analysis while taking account of their correlation structure, we also used analyses of variance with repeated observations (mixed models). In these models the siblings were the repeated events. Data handling and statistical analyses were performed using SPSS version 11.0 software (24).
![]() |
RESULTS |
---|
![]() ![]() ![]() ![]() ![]() ![]() ![]() |
---|
Analyses of gestational age gave a similar pattern of results. The malformed infants had on average 47 days' shorter gestations than did infants of the same birth order in unaffected families, whereas the mean gestational ages of the malformed infants' siblings were close to those of their counterparts in unaffected families.
We used standard analyses of variance to adjust the birth weight differences between infants of the same birth order in unaffected and affected families for gestational age. This reduced the birth weight differences for the malformed infants but had only little impact on the birth weight differences between nonmalformed siblings and corresponding control infants (table 2).
Further, we evaluated confounding by the mother's age, the mother's educational level, maternal diabetes, marital status, time period of first birth, and interpregnancy interval (table 2). The estimated birth weight differences did not change notably by including these variables in the models. The intercepts changed in an expected manner according to the different reference groups. The proportion of preeclamptic pregnancies was not significantly different in families with and without a malformed infant, and preeclampsia was thus not included in the multivariable analysis.
In families with at least three infants, the results were similar (figure 1). The crude mean birth weight was significantly reduced for infants with a registered congenital malformation, whereas their nonmalformed siblings had values almost equal to those of infants with the same birth order in unaffected families. This was found for siblings born both before and after the birth of a malformed infant and also for the nonmalformed siblings in families where two of the three infants had a birth defect.
|
|
When the malformed infant was registered with a neural tube defect, an isolated cleft palate, a cardiac defect, limb defect, abdominal wall defect, external genital defect, or Down's syndrome, the nonmalformed siblings' crude mean birth weights did not differ significantly from those of corresponding control infants. A statistically reduced mean birth weight was found for siblings of infants with multiple malformations and for second-born infants in sibships where the first infant was registered with a cleft lip. This was so also after adjusting for the sibling's gestational age, mother's age, mother's education, maternal diabetes, marital status, interpregnancy interval, and time period (table 4).
The sex ratio (boys/girls) was significantly higher among infants with external genital defects, cleft lip, and limb defects than among control infants, but this was not the case among the nonmalformed siblings in any of the malformation categories.
Table 4 displays considerable variability in the reduction of mean birth weight for infants with different types of malformations. The following malformation categories were associated with a reduction in mean birth weight of at least 1,000 g compared with control infants of the same parity: neural tube defects, hydrocephalus, other central nervous system, and other syndromes. We grouped these into one "severe birth weight reduction" category and analyzed the nonmalformed siblings' crude and adjusted mean birth weight, as before. First-born siblings (born before a malformed infant in this group) had a 33-g lower mean birth weight than did corresponding control infants (95 percent confidence interval (CI): -76, 10), and second-born siblings (born after a malformed infant) had a 44-g lower mean birth weight (95 percent CI: 83, 6). When adjustment was made for gestational age, mother's age, mother's educational level, maternal diabetes, marital status, time period of first birth, and interpregnancy interval, the estimated differences were 7 g higher (95 percent CI: 31, 45) and 22 g higher (95 percent CI: 13, 57) than control infants, respectively.
![]() |
DISCUSSION |
---|
![]() ![]() ![]() ![]() ![]() ![]() ![]() |
---|
It is well established that there is a strong relation between the birth weights of siblings and that growth restriction and preterm delivery recur in sibships (38
). The correlation coefficient for birth weights of two successive siblings is in the range of 0.5, and a woman's risk of delivering a low birth weight infant in her second birth is 510 times higher for women whose first infant was low birth weight than for women whose first infant was not low birth weight (3
5
, 25
, 26
).
Infants with birth defects are more often found to be growth restricted than are infants without such defects, the magnitude of growth restriction varying between types of malformations (1, 2
). In the study by Khoury et al. (1
), only a few isolated defects (such as isolated polydactyli, pyloric stenosis, and congenital hip dislocation) were not associated with excessive intrauterine growth restriction. The relation between growth restriction and congenital malformations may be explained by either growth restriction as the primary problem, predisposing the fetus to malformations (27
), or growth restriction as the result of, or being a reaction to, the presence of malformations (28
). The presence of underlying factors related to both growth restriction and congenital malformations may also explain the association (29
, 30
). These mechanisms are not necessarily mutually exclusive, and they may operate differently for different types of malformations.
Spiers suggested that "malformations will be more likely to occur when the overall growth rate of the fetus has already been grossly disturbed" (27, p. 312) and that "overall growth retardation constitutes a state of increased susceptibility to congenital malformations" (27
, p. 314). In a study of early fetal growth with ultrasound measures of 99 pregnant diabetic women, Pedersen and Molsted-Pedersen (29
) found that seven of nine fetuses with congenital malformations were growth delayed at the time of ultrasound measurement (from the seventh to 14th week of gestation). They suggested a common mechanism behind early growth delay and induction of abnormal embryogenesis.
The results of our study show that reduced birth weight associated with congenital malformation is restricted to the pregnancy with the malformed fetus. Unlike other situations where growth restriction is found, there seems to be little or no relation between the malformed infant's growth and the growth of its siblings. If growth restriction were one of the primary mechanisms behind congenital malformations, one would expect to find similar sibling relations for birth weight in families with a growth-restricted malformed infant as for families with a growth-restricted nonmalformed infant. Instead, we find that, in families with a malformed and growth-restricted infant, nonmalformed siblings' birth weights are hardly at all affected by the fact that one of the siblings is growth restricted.
This may also suggest that persisting factors (biologic, environmental, or socioeconomic) play different roles for the growth restriction associated with congenital anomalies than for growth restriction not associated with such. Short maternal height is an example of a persistent biologic factor associated with lower birth weight among offspring (12) but not with risk of birth defects. Smoking in pregnancy is an acknowledged risk factor for fetal growth restriction (12
, 31
, 32
), but there has been much controversy about what role smoking plays in the etiology of birth defects (32
). Low socioeconomic status is also found to be associated with lower birth weight (12
, 20
, 33
), whereas the association between social status and birth defects is more unclear (34
).
Subgroup analyses
Cleft lip and multiple defects were the only subgroups where we found significantly lower birth weight among nonmalformed siblings than among the corresponding control infants.
As mentioned above, smoking is associated with growth restriction and may represent a "persistent factor" as mothers who smoke in one pregnancy are likely to smoke in all. Several studies have been conducted on the association between smoking in pregnancy and risk of cleft lip and/or palate in the offspring, including investigations of gene-environment interactions. Results are inconsistent, but many find a moderately increased risk of cleft lip and/or palate among the offspring of smokers (32, 35
, 36
).
Socioeconomic status will also often be a persistent factor across a woman's pregnancies. A recent study (34) has found an increased risk of multiple malformations among mothers from lower socioeconomic levels than from higher, and our finding of lower birth weight among nonmalformed siblings in a family with one multiply malformed infant agrees with this. Multiple malformations are also a group that is important for detecting human teratogens, as most potent teratogens are associated with multiple rather than isolated malformations (37
). Teratogens may be limited to one pregnancy, but they may also represent a persistent factor that influences more than one pregnancy if, for instance, present in the mother's environment. Intrauterine growth restriction has also been suggested as a manifestation of teratogenic action in humans and as a possible endpoint in mutation epidemiology (38
). Environmental teratogens are most likely associated with socioeconomic status.
When combining the malformation categories associated with the most severe reductions in mean birth weight (1,000 g or more), we found that the main results for nonmalformed siblings were still the same: Their crude and adjusted mean birth weights did not differ significantly from those of control infants. The expected crude mean birth weight for second siblings following an infant that weighs 1,000 g less than the "population mean," but without having a malformation, is 3,100 g. This is 470 g less than the population mean weight of second-born infants. In contrast, second siblings born after a malformed infant with an average birth weight reduction of 1,000 g weighed only 44 g less than corresponding control infants.
Problems with the study
The Medical Birth Registry of Norway is a compulsory notification system, which has been running since 1967. The unique national identification number ensures identification and linkage between registries. The registration of births after 16 weeks of gestation should be considered complete except for abortions shortly after week 16.
One of the main problems with the present study is the ascertainment of cases. Incomplete ascertainment of congenital anomalies in registries is an acknowledged problem (39) and has been discussed also for specific birth defects in the Norwegian Birth Registry (40
). The registered defects are based on diagnoses made during the infants' stay at the maternity ward, and some malformations (for instance, certain cardiac and some urinary defects) will not be diagnosed until later. Low ascertainment means that there may be infants with undiagnosed malformations among the nonmalformed siblings and among the controls. However, whereas there may be higher ascertainment among siblings following the birth of a malformed infant, there is no reason to believe that there are more undiagnosed malformations among the control infants than among the infants delivered preceding a malformed baby. In these cases the low ascertainment will affect the controls and the nonmalformed siblings equally, and the difference found should not be biased.
Infants with multiple malformations may also mistakenly be classified as having a single defect. Our findings indicated that siblings of multiply malformed infants had lower mean birth weights than did control infants, whereas siblings of single-defected infants did not differ from control infants. The mentioned misclassification would lead the birth weight differences away from the null for siblings of single-defected infants and probably toward the null for the siblings in the multiple malformations group. In other words, if this misclassification is sizable in our data, the "true" results will probably be even more pronounced in the direction that we have already described.
Our subgroup analyses were based on rather broad categories, and it is possible that a more detailed classification would give results that deviate from the general pattern we report, this being true also for other subgroups of malformations. In addition, the group of multiple malformations is heterogeneous, and again, a more detailed analysis may disclose different patterns within this group. However, the chances of finding significant results when more homogeneous categories are analyzed diminish because of smaller numbers.
Some risk factors that are known or suspected to affect birth weight were not included in the multivariable analyses because of lack of information. One of the most important factors is maternal smoking. However, social status and smoking are correlated variables, and some researchers mean that the main pathway by which social status affects birth weight is through smoking (41). We may thus have adjusted for the smoking variable indirectly by adjusting for social status. The mother's height and prepregnant weight are also known to affect birth weight and are variables not contained in the Medical Birth Registry. To be a confounder, the variable must be associated with both the outcome and the exposure, in this case with both birth weight and being a sibling of a malformed infant. However, we do not believe that there are evident reasons to believe that maternal height and prepregnant weight are associated with an increased risk of giving birth to a malformed infant by pathways that are not contained in the social status of the mother.
Evaluation of risk
Compared with women with nonmalformed first infants, women who have delivered a first-born baby with a congenital anomaly have a 2.4 times higher risk of delivering a second infant with any registered defect, mainly because of an increased risk of the same type of defect (21). Among women delivering their first-born babies, 2.5 percent gave birth to infants with any type of birth defect, and 2.1 percent delivered a malformed second infant after a healthy first birth.
Eugenic theory in the beginning of the 20th century viewed congenital malformations as an expression of "physical degeneracy" that could be traced to polygenically poor protoplasm. Galton in 1909 and Pearson in 1907 viewed the decline of a population to be caused by the heredity of continuously variable protoplasm by entire groups in society and, for instance, harelip and cleft palate were simply the manifestation of polygenically inherited physical degeneracy by certain groups (42). Our study shows no indication of "physical degeneracy" in the families where one or more of the infants are born with a congenital malformation.
We conclude that women who have delivered an infant with a congenital anomaly should be reassured that, apart from the increased risk of a similar anomaly in the next birth (which in absolute terms is small), there seems to be little reason to fear an adverse birth outcome associated with growth restriction in her next pregnancy.
![]() |
ACKNOWLEDGMENTS |
---|
The authors thank Dr.Tine Brink Henriksen for valuable comments.
![]() |
NOTES |
---|
![]() |
REFERENCES |
---|
![]() ![]() ![]() ![]() ![]() ![]() ![]() |
---|