The Risk of Low Birthweight VICTOR EISNER, MD, MPH, JOSEPH V. BRAZIE, MD, MPH, MARGARET W. PRATT, AB, AND ALFRED C. HEXTER, PHD
Abstract: Descriptive analyses of birthweight for single live births in the United States during 1974, using birth certificate information, show that several factors are associated with a high incidence of low birthweight babies. Multivariate analyses were performed to determine correlates of low birthweight. When other factors are held constant, race not white, previous reproductive loss, short interpregnancy interval, outof-wedlock birth, no prenatal care, and maternal age
under 18 years or over 35 years each increase the risk of having an infant of low birthweight. High birth order and maternal education under 12 years of schooling were not consistent risk factors for low birthweight. Odds ratios are presented and the method for combining these, to estimate the risk for an individual mother having a low-birthweight infant, is illustrated. (Am J Public Health 69:887-893, 1979.)
Recent studies of pregnancy outcome have shown a progressive decline of perinatal and infant mortality in all risk categories, including mortality for small preterm infants.' The decline has come at a time of change in many aspects of maternal and infant care: associated factors include improved perinatal services available to both mother and baby, improved care or prevention of specific diseases, and an increased availability of means for spacing or preventing pregnancies. In our study of birth certificate information to evaluate specific factors which increase the risk of a poor pregnancy outcome, we have followed the suggestion of Lewis, et al.,2 and use birthweight distribution as an indicator of pregnancy outcome. Although infant mortality rates remain the primary indicator, low birthweight is also an indicator of considerable interest. Low birthweight is associated with increased perinatal and infant mortality and morbidity, including adverse sequelae such as mental retardation and learning disabilities. Many previous studies have shown that birthweight, in-
fant mortality, and infant morbidity vary with such strongly interacting factors as maternal age, social class, previous reproductive history, and prenatal care.3-'0 Most studies, however, have not attempted to determine the specific effect of single factors. Gendell and Hellegers"I in a linked-records study were able to attribute one-fourth of the drop in Baltimore perinatal mortality over a five-year period to an increase in lower-order births and younger mothers. Kessner, et al. 10 attempted to isolate the effects of health care on infant mortality by analyzing separate groups divided by social and medical risk of high mortality. They concluded that within any risk group, mortality varied inversely with the amount of care received. Dott and Fort,'2 also using a linked-records study, held socioeconomic status and birthweight constant and showed that infant mortality varied inversely with the number of prenatal visits. Finally, Morris, Udry and Chase'3 used an adjusted-rate method to attribute 27 per cent of the fall of infant mortality from 1965 to 1972 to changes in age and parity of the mothers. The study from which the present report has been drawn'4 comprised an examination of United States birth registration data for single live births in 1974. For each race it related the proportion of births which were low birthweight (less than 2501 grams) to risk factors ascertainable from birth certificates. We summarize here the principal findings of that study and report the results of multivariate analyses which examine the risk of having a low birthweight baby by race, maternal age, wedlock status, prenatal care, and maternal education for primigravida, and by these factors and birth order, reproductive history, and interpregnancy interval for multigravida.
Address reprint requests to Joseph V Brazie, MD, MPH, Maternal and Child Health, Annex 4, Room 500, California Department of Health Services, 2151 Berkeley Way, Berkeley, CA 94704. Dr. Eisner is with the San Francisco Department of Public Health; Ms. Pratt is with the Information Sciences Research Institute, Silver Spring, MD; and Dr. Hexter is with the University of California School of Public Health, Berkeley. This paper, submitted to the Journal February 12, 1979, was revised and accepted for publication May 17, 1979. Editor's Note: See also editorials pages 851, 852.
A-JPH September 1979, Vol. 69, No. 9
887
EISNER, ET AL
Methods The study universe consisted of all single live births resident and recorded in the 50 United States and the District of Columbia during 1974. The original data file, obtained from the National Center for Health Statistics, included all records from 16 states and a 50 per cent sample from the remainder. Thus, the basic file consisted of 1,989,128 records representing 3,101,117 births. Only 30 states reported all variables, and not all of the certificates for these states provided all information requested. Therefore, the multivariate analyses were based on a subset of the basic file consisting of 505,243 records with complete information on race, age of mother, onset of prenatal care, total birth order, maternal education, wedlock status and, for multigravida, interpregnancy interval and reproductive history.* Analyses were done by the Maternal and Child Health Studies Project of the Information Sciences Research Institute, Silver Spring, MD, with further analyses by the Collaborative Research Project of the Maternal and Child Health Program, School of Public Health, University of California, Berkeley. The ideal dependent variable for the multivariate analyses would be a combination of birthweight and gestational age.'5 However, the analyses for the descriptive portion of our study raised serious questions about the reliability of gestational age reporting. For example, 22 per cent of the births with reported gestational age less than 26 weeks weighed more than 2500 grams. Numerous studies, e.g., Lubchenko,'6 have shown that no births with true gestational age less than 26 weeks weigh more than 2500 grams. It appeared to us that the majority of the errors were in gestational age rather than birthweight and were concentrated in the group of greatest interest to us.** Preliminary analyses, using a combination of birthweight and gestational age as dependent variable, gave essentially the same results as those using birthweight alone. We concluded that, for our analyses, a dependent variable combining gestational age and birthweight gave no advantage over one using birthweight alone. *Comparison of marginal tabulations for the basic file with the file used for multivariate analyses showed small but statistically significant differences. The multivariate file had more births weighing less than 2501 grams, fewer with maternal age under 18 years, maternal education under 12 years of schooling, no prenatal visits or birth out-of-wedlock, and more with previous losses or interpregnancy interval less than six months. While we do not believe these affect our conclusions, the possibility that births with complete information are not representative of the universe of all births cannot be excluded. **The National Center for Health Statistics calculates gestational age from date of birth and date of last menstrual period (LMP). This procedure is much more reliable than the previous method of asking directly the number of completed weeks of gestation. However, Williams17 has shown that this procedure tends to misclassify certain births, especially those with very short interpregnancy intervals where LMP may refer to the previous birth. The number misclassified is small but enough to bias seriously the lower gestational age groups. 888
We defined two birthweight groups as dependent variables for the multivariate analyses: 1. birthweight less than 1501 grams, all gestations 2. birthweight less than 2501 grams, all gestations These two birthweight groups are not independent because all births less than 1501 grams are also included in births less than 2501 grams. However, only about 15 per cent of the births under 2501 grams are also under 1501 grams, and analyses of the two groups showed substantial differences. Our preliminary studies treated race as an independent variable which could take on different values in a single analysis. However, Black mothers were more likely than White mothers to have low birthweight babies at every level of every factor studied. For this reason we made separate analyses of each racial group (White, Black and Other) for primigravida and for multigravida. Regression-type analyses with more than 500,000 observations were beyond our resources; therefore we grouped the observations. The groupings used for the independent variables were: 1. Maternal Age: Under 18 years, 18-34 years, 35 years and over; 2. Wedlock Status: In-wedlock, Out-of-wedlock; 3. Prenatal Care: First trimester, Second trimester, Third trimester, No care (in some analyses these were grouped into "Some care" and "No care"); 4. Maternal Education: Under 12 years of schooling, 12 years of schooling and over; 5. Birth Order (for multigravida): Less than six, six and over; 6. Reproductive History (for multigravida): No previous reproductive losses, one or more previous losses; 7. Interpregnancy Interval (for multigravida): Less than six months, six months and over; These categories provided 1,296 possible combinations of independent variables (144 for primigravida, 1,152 for multigravida), of which 390 contained no births. Thus, we had 906 "data points" for the multivariate analyses. Initial screening showed only minor differences in per cent low birthweight for the three trimesters of onset of prenatal care, so these were grouped into "Some care" vs. "No care" for reporting the multivariate comparisons (Table 1). We used two methods to estimate the relative importance of each independent variable: multiple logistic regression (MLR), and the Mantel-Haenszel (M-H) procedure. MLR'8. '9 is a linear regression which uses the logarithm of the relative proportion of cases to non-cases as the dependent variable. It is closely related to discriminant analysis, and may be regarded as an extension of discriminant analysis to a broader class of models.20 21 The regression coefficient for each independent variable provides an estimate of the "odds ratio," the ratio between the chance an event will occur when a given factor is present to the chance it will occur when the factor is absent. For example, an odds ratio of 4.95 for the risk factor "No prenatal care" with respect to birthweight less than 2501 grams can be interpreted to mean that if all other factors are the same a woman who has not received prenatal care is about five times as likely to give birth AJPH September 1979, Vol. 69, No. 9
RISK OF LOW BIRTHWEIGHT
to an infant weighing less than 2501 grams as is a woman who has received some prenatal care. Similarly, an odds ratio less than 1.0 means the risk is reduced, rather than increased, if the factor is present. The M-H procedure22 23 is a chi-square-based method for combining contingency tables to assess the significance of a particular classifying variable in the presence of identified nuisance variables. It provides a chi-square with one degree of freedom, expressing whether low birthweight appears significantly more (or less) often in cases with the factor than in cases without the factor being studied. It also provides an estimate of the odds ratio, which may differ from the estimate provided by MLR. Because of time and budget limitations we did not test all categories of variables by both procedures.
of-wedlock being 6.4 per cent for Whites, 47.5 per cent for Blacks and 14.4 per cent for Others. Substantially higher rates for low birthweight babies were observed for out-ofwedlock births, for all races, with the proportionately greatest increase for Whites (Table 1). The M-H procedure, which was done for primigravida, showed a significantly increased risk of low birthweight for White and Black primigravida born out-of-wedlock but not for Other primigravida. Odds ratios for Whites and Blacks ranged from 1.17 to 1.45. Odds ratios by MLR for primigravida were similar. Odds ratios by MLR for multigravida were substantially higher than for primigravida for White and Other out-of-wedlock births (1.54 to 3.41) but not for Blacks (Table 3). Prenatal Care
Results The distribution of births and of the per cent of births less than 2501 grams are shown for each independent variable in Table 1. In considering the results it should be borne in mind that the "Other" race group is quite heterogeneous and comprised only about 2.5 per cent of all births. Therefore, while results are presented for the "Other" race group, with statements concerning statistical significance, the results for this group should be considered cautiously. The effectiveness of the MLR can be summarized by multiple correlation coefficients. In general, the regressions "explained" more of the variance for the primigravida than for the multigravida, for the births less than 1501 grams than for the births less than 2501 grams, and for the Whites than for the Blacks or Others (Table 2). The results of the M-H procedures and of the MLR are shown in Table 3. Maternal Age Black mothers were more likely to be under 18 years of age than White or Other mothers. Thus, Black mothers had substantially more births in the age group with the highest proportion of low birthweight babies (Table 1). The MLR showed that for all three races primigravida over 35 years of age had substantially increased risk of low birthweight compared to mothers aged 18-34 years; odds ratios ranged from 2.00 to 8.86. Multigravida over 35 years of age also had increased risk but not so great as for primigravida. The M-H procedure was not done for this variable (Table
3).
Maternal age under 18 years was a substantial risk factor for multigravida, with odds ratios ranging from 1.34 to 6.72. Young maternal age was not an important factor for primigravida when other factors were held constant; odds ratios ranged from not significantly different from 1.0 to a maximum of 2.74. Wedlock Status A substantial difference for proportion of births by wedlock status is evident between races, with the per cent outAJPH September 1979, Vol. 69, No. 9
Black mothers tended to have their first visit for prenatal medical care somewhat later than did White or Other mothers. However, for all three race groups the median time, for those with time of first visit reported, was in the first trimester. The time of first prenatal visit made very little difference in per cent of births which were low birthweight, provided the mother had at least one prenatal visit. However, no prenatal medical care was a very significant risk factor for the 1.4 per cent of mothers who lacked it. No prenatal care was the greatest risk factor for low birthweight in our study. The M-H procedure showed highly significant increased risks of birthweights below 1501 grams and below 2501 grams for primigravida, and for mothers with no prenatal care in all three race groups; odds ratios ranged from 2.78 to 15.65. Odds ratios from MLR were similar to those from M-H, for primigravida. Odds ratios by MLR for multigravida with no prenatal medical care were statistically significant, ranging from 2.09 to 5.88. In every case, the odds ratios for multigravida were substantially lower than for the corresponding group of primigravida. The M-H procedure was not done for multigravida, for this variable (Table 3).
Maternal Education Most mothers had achieved 12 years of education (high school graduation) or more: 74.5 per cent for Whites, 53.6 per cent for Blacks, and 67.1 per cent for Others. Achievement of at least high school graduation was associated with improved outcome in terms of low birthweight rates (Table 1).
The multivariate analyses gave paradoxical results for this factor. The M-H procedure showed strong associations between maternal education under 12 years of schooling and birthweights below both 1501 grams and 2501 grams for White primigravida and for birthweights below 2501 grams for Black primigravida but not for Other primigravida. The same pattern occurred for multigravida. The MLR, however, gave quite different results. Of the six odds ratios calculated for primigravida, only that for White primigravida was statistically significant. For multigravida four of the six odds ratios were statistically significant, and all four were in the opposite direction to that expected (Table 3). 889
EISNER, ET AL TABLE 1 -Number of Births and Per Cent Le than 2501 Grams for Selected Variables by Race for Single Live Births, United States, 1974 Number of Births
Variable
Total
TOTAL 3,101,117 Maternal Age (years) Less than 15 12,436 15-17 231,691 18-19 356,461 20-24 1,089,049 25-29 903,573 30-34 363,872 35-39 115,080 40-44 27,260 45 and over 1,695 Wedlock Status In-wedlock 1,812,642 Out-of-wedlock 280,682 Not stated 10,570 Not reported' 997,223 Prenatal Care First trimester 1,852,701 Second trimester 557,092 Third trimester 123,935 No prenatal care 35,871 Not stated 109,994 Not reported' 421,524 Matemal Education (years) 0-8 134,794 9-11 508,463 12 994,045 13-15 325,635 16 and over 239,482 Not stated 59,172 Not reported' 839,526 Birth Order First 1,220,082 Second 926,821 Third 451,258 Fourth 214,679 Fifth 105,515 Sixth 55,559 Seventh 31,698 Eighth and over 48,016 Not stated 47,489 Reproductive History No previous losses 1,311,558 One or more previous losses 405,902 No previous pregnancy 1,146,894 Not stated 236,763 Interpregnancy Interval Less than 6 months 116,514 6-11 months 133,685 12-23 months 263,377 24 months and over 605,387 No previous pregnancy 964,762 Not stated 465,106 Not reported' 552,286
Per Cent Less than 2501 Grams
White
Black
Other
Total
White
Black
Other
2,529,411
495,950
75,756
6.5
5.4
12.0
6.5
5,010 150,777 264,655 896,737 786,497 309,583 93,248 21,591 1,313
7,241 76,987 85,009 169,925 92,467 41,569 17,610 4,802
185
3,927 6,797 22,387 24,609 12,720 4,222
15.3 10.5 8.4 6.2 5.2 5.8 7.1 8.4 9.6
12.2 8.5 6.9 5.2 4.6 5.2 6.3 7.5 9.4
17.5 14.5 13.3 11.4 10.1 10.6 11.5 12.2 10.6
9.2 8.3 8.0 6.4 5.8 6.2 7.8 10.0 7.1
1,590,022 108,050 9,109 822,230
183,240 166,031 1,285 145,394
29,599
5.8 11.9 6.9 6.3
5.2 9.1 6.1 5.4
10.4 13.9 12.1 11.9
6.4 8.4 9.1 6.3
1,592,762 400,029
217,871 140,049 34,324 12,412 25,404 65,890
42,068 17,014 5,214 1,389 1,697 8,374
5.7 7.7 7.8 18.9 8.7 6.7
4.9 6.2 6.4 15.3 6.8 5.7
11.0 12.1 11.5 25.7 14.7 11.8
6.2 6.8 5.7 14.5 8.2 6.5
29,810 143,084
5,275 11,132
147,812 36,187 15,430 13,639 109,988
16,667 7,134 9,640 1,238 24,670
9.3 9.4 5.9 5.0 4.2 7.8 6.2
8.0 7.7 5.0 4.3 3.8 6.0 5.4
14.0 13.6 11.0 10.1 8.7 13.7 11.6
7.2 7.4 6.5 5.9 6.0 6.5 6.4
775,851 369,776 171,143 81,853 41,643 22,963 31,999 36,588
193,015 128,104 70,360 38,466 21,012 12,461 7,886 14,569 10,077
29,472 22,866 11,122 5,070 2,650 1,455 824
6.9 5.8 6.1 6.8 7.3 7.9 8.0 8.4 6.8
5.9 4.8 5.0 5.6 6.3 6.8 6.7 7.1 5.3
12.5 11.7 11.6 11.9 11.5 11.6 11.5 11.4 12.3
6.9 6.2 6.0 6.5 6.5 7.5 8.1 6.6 6.8
1,083,255
195,368
32,935
5.4
4.5
10.4
5.5
324,500
70,707
10,695
9.0
7.6
15.2
8.7
940,840 180,816
177,671 52,204
28,383
3,743
6.9 6.7
5.9 5.1
12.5 12.1
7.0 6.0
92,979 108,287 223,994
19,341 20,832 32,102
4,194 4,566 7,281
7.8 6.3 5.1
6.4 5.2 4.3
14.5 12.1 10.7
8.1 6.7 5.7
502,851
88,961
13,575
5.2
4.4
9.7
4.7
793,079 355,287 452,934
145,003 97,321 92,390
26,680 12,498 6,962
5.8 6.3 5.8
12.5 13.3
6.9 7.6 6.9
84,397 22,070 82,893 347,260
99,709 354,247 829,566 282,314 214,412 44,295 704,868
997,595
340
867 42
39,380 6,601 176
849
1,448
6.8
7.8 6.8
11.9
'This item not reported on birth certificates in a number of states.
890
AJPH September 1979, Vol. 69, No. 9
RISK OF LOW BIRTHWEIGHT
TABLE 2-Multiple Correlations (R2) for Multiple Logistic Regressions Birthweight Group Race and Gravidity
Primagravida White Black Other Multigravida White Black Other
Less than 1501 grams
Less than 2501 grams
.863 .725 .663
.770 .512 .679
.600 .540 .558
.422 .377 .322
Birth Order
Primigravida comprised about 40 per cent of the births in all racial groups. The numbers decreased rapidly with increasing birth order, more rapidly for Whites than for the other two race groups; 7.2 per cent of the Black births were birth order six or higher, compared to 3.9 per cent for Whites and 5.0 per cent for Others. For Whites and Others the lowest per cent low birthweight was observed for birth orders two to four (two to five for Others). For Blacks the low birthweight per cent was 12.5 per cent for first births, and lower for the higher order births (Table 1). The effect of high birth order (gravida six and over) was examined using MLR. When other variables were taken into account the effect of high birth order was small (Table 3). Reproductive History Percentages of multigravida with one or more previous losses prior to this delivery were 23.1 per cent for Whites, 26.6 per cent for Blacks, and 24.5 per cent for Others. The association of reproductive history with the birthweight outcome of the current pregnancy is striking. Each race showed about a 50 per cent increase in per cent low birthweight for one or more previous losses, compared to no previous losses. The low birthweight rates for primigravida were intermediate (Table 1). The M-H procedure indicated a statistically highly significant increased risk of low birthweight for mothers with one or more previous losses for all three race groups, for both birthweight categories; the odds ratios ranged from 1.56 to 2.69. Similar results were obtained by the MLR, with odds ratios of 1.34 to 2.22 (Table 3).
Interpregnancy Interval The per cent of multigravida mothers with interpregnancy interval less than six months was 10.0 per cent for Whites, 12.0 per cent for Blacks, and 14.2 per cent for Others. A short interpregnancy interval was associated with an increased per cent of low birthweight babies. The lowest proportions of low birthweight babies occurred to mothers who waited at least 12 months after the previous delivery before again becoming pregnant (Table 1). The M-H procedure indicated an increased risk of low birthweight for both birthweight categories to White and AJPH September 1979, Vol. 69, No. 9
Black multigravida with interpregnancy intervals of less than six months; odds ratios ranged from 1.12 to 1.55. The odds ratio for Other multigravida was substantial for births less than 1501 grams (1.81) but was not significantly different from 1.0 for births less than 2501 grams. MLR indicated substantially increased risk of low birthweight for all three race groups and both birthweight categories; odds ratios by MLR ranged from 1.31 to 2.69 and in every case were higher than the odds ratios by the M-H procedure (Table 3).
Discussion and Conclusions This study supports the findings of previous studies which associate a poor outcome of pregnancy with certain risk factors. It also permits certain conclusions for program planning. Both the regression (MLR) analyses and the MantelHaenszel (M-H) procedure show that a previous pregnancy loss, an interpregnancy interval less than six months, a birth out-of-wedlock, and a pregnancy without prenatal medical care all increase the risk of having a low birthweight baby, independent of each other and of the other variables studied. The magnitude of the increased risk varies with gravidity and race. In general the greatest risk comes from pregnancies with no prenatal medical care, and the risks from the other factors, while appreciable, are considerably smaller. The regression analyses of maternal age show a higher risk for multigravida under 18 years of age than for primigravida in this age group, and a higher risk for primigravida over 35 years of age than for similarly aged multigravida. These findings, and the inconsistently higher risks for the sixth and higher birth orders, conform with conventional wisdom. The two methods of analysis showed contradictory results for the effect of maternal education. This occurred because the two forms of multivariate analysis measure different things. The M-H procedure, like most chi-squarebased tests, is primarily reporting the largest groups. MLR, which looks only at the proportions, like most regression procedures gives greatest weight to observations furthest from the mean and therefore is primarily reporting consistency. From a public health standpoint the results of M-H, which showed low maternal education to be a risk factor for Whites for both birthweight groups, and for Blacks for birthweights below 2501 grams, would appear to be more meaningful. The regression results indicating a contrary effect suggest that social and environmental effects are operating which were not identified by birth certificate information and are not randomly distributed over the categories used. However, the odds ratios derived from maternal education are small. If this factor has an independent effect, the effect is of little importance. This study indicates that lack of prenatal care is an important risk indicator for low birthweight. Lack of prenatal care, however, not only indicates the possibility that needed medical intervention did not occur, it also may indicate nonconventional health behavior, including maternal behaviors directly detrimental to the infant. Also, an unknown proportion of births with "no care" represent premature deliveries 891
EISNER, ET AL TABLE 3-Results of Mantel-Haenszel Procedures and Multiple
Logistic Regressions Odds Ratio
Chi-square' from Mantel-Haenszel Risk Factor, Race, and Gravidity
Birth less than 1501 grams
Maternal age less than 18 years White primigravida Black primigravida Other primigravida White multigravida Black multigravida Other multigravida Maternal age 35 years and over White primigravida Black primigravida Other primigravida White multigravida Black multigravida Other multigravida Birth out-of-wedlock White primigravida 29.33** Black primigravida 5.88* Other primigravida 0.33 White multigravida Black multigravida Other multigravida No prenatal care White primigravida 710.27** Black primigravida 361.36** Other primigravida 42.53**
White multigravida
Black multigravida Other multigravida Maternal education less than 12 years White primigravida 7.03** Black primigravida 2.82 Other primigravida 0.16 White multigravida 28.41" Black multigravida 0.72 Other multigravida 0.18 Birth order six and over White multigravida Black muftigravida Other muftigravida One or more previous pregnancy losses White multigravida 226.04** Black muftigravida 125.21** Other multigravida 15.32** Interpregnancy interval less than six months White multigravida 21.09** Black multigravida 19.14** Other multigravida 4.49*
Birth less than 2501 grams
63.18** 15.78** 0.04
from Mantel-Haenszel2 Birth less than 1501 grams
1.45** 1.23* 1.38
Birth less than 2501 grams
1.29** 1.1 7** 1.04
from regressions Birth less than 1501 grams
2.71
4.10** 5.77** 8.86** 2.45** 2.19** 2.63**
2.00" 3.56" 1.27" 1.17 1.70"
1.41 *
1.12
1.58*
1.52' 1.09 1.76" 1.28" 1.54" 4.95*' 2.83" 5.91" 2.92" 2.09" 2.75"
1.53 1.24*
2.25" 11.53** 6.19** 15.65**
4.92** 2.78** 4.80**
14.97" 5.05" 14.29" 5.88" 3.45" 4.04"
117.64** 23.52** 0.01 491.22** 12.89** 0.44
1.20** 1.17 1.27 1.38** 0.92 0.86
1.36** 1.24** 1.02 1.71 ** 1.13** 0.93
1.21 0.96
1.79** 1.16 2.74** 4.38** 2.93** 6.72**
3.41" 451.67** 179.08** 20.83**
Birth less than 2501 grams
1.07 1.12 0.95
0.72" 0.52"
1.65' 1.63" 1.34"
2.11
1.32* 0.97 0.88 1.01
0.84*
0.73*
0.82
1.58"
0.98 0.85 1.38*
0.97
2.21" 484.04** 175.62** 14.98**
2.33** 2.61 ** 2.69**
1.63** 1.66** 1.56**
1.63" 2.22" 1.52"
1.54" 1.71"
15.91 **
1.38** 1.55** 1.81*
1.12** 1.28** 1.29
2.33" 2.69" 2.12"
1.31" 1.60" 1.57"
31.44** 3.37
1.34*
'All chi-squares have one degree of freedom. 2Significance levels for odds ratios from Mantel-Haenszel are by convention taken to be the same as for the associated chi-square.
*.01 < p < .05 **p < .01
to women who would have obtained care in the third trimester had the pregnancy not ended. This study showed that lack of prenatal care is an indicator of risk but does not allow conclusions about the effect of providing earlier care. In the same way, measures to increase interpregnancy intervals may contribute to substantial reductions in risk but our study 892
does not provide direct proof of this. The study shows that women with a previous pregnancy loss remain at high risk independent of their age, race, education, interpregnancy interval, gravidity, or wedlock status. While primigravid women pregnant out-of-wedlock are not necessarily at high risk, a woman who has a second (or higher) pregnancy out-of-wedAJPH September 1979, Vol. 69, No. 9
RISK OF LOW BIRTHWEIGHT
lock has a markedly increased risk of a low birthweight delivery. The expression of risks as odds ratios allows an estimate of individual risk by multiplying together the odds ratios for each risk factor borne by the individual. For example, if we assume risk factors are independent and the MLR odds ratios from this study apply, the risk of having an infant under 2501 grams for an unmarried White multigravida over 35 years of age, with a previous loss, can be estimated as 1.76 x 1.27 x 1.54 = 3.44 times the risk for a White multigravida who is married, 18-34 years of age, with no previous losses, all other factors being equal. In terms of total contribution to the number of infants with low birthweight in our study, 401 White unmarried multigravida over 35 years of age, with a previous loss, had 40 low birthweight infants compared to the 6,505 low birthweight infants delivered to the 164,805 White married multigravida 18-34 years of age, with no previous loss. Thus, this high risk group actually made a very small contribution to the overall problem, even though individually they were at substantially higher risk.
REFERENCES
1. Eisner V, Pratt MW, Hexter A, et al: Improvement in infant and perinatal mortality in the United States, 1%5-1973: I. Priorities for intervention. Am J Public Health 68:359-364, 1978. 2. Lewis R, Charles M, Patwary KM: Relationships between birth weight and selected social, environmental and medical care factors. Am J Public Health 63:973-981, 1973. 3. Shapiro S, Unger J: Weight at Birth and Its Effect on Survival of the Newborn in the United States, Early 1950. Vital StatisticsSpecial Reports, Vol. 39 No. 1. U.S. National Office of Vital Statistics, Public Health Service, Washington, DC, 1954. 4. Shapiro S, Unger J: Relation of Weight at Birth to Cause of Death and Age at Death in the Neonatal Period: United States, Early 1950. Vital Statistics-Special Reports, Vol. 39 No. 6. U.S. National Office of Vital Statistics, Public Health Service, Washington, DC, 1956. 5. Butler NR, Bonham DG: Perinatal Mortality: The First Report of the British Perinatal Mortality Survey. Edinburgh: E & S Livingstone, Ltd., 1%3. 6. Butler NR, Alberman ED, (Eds.): Perinatal Problems: The Second Report of the British Perinatal Mortality Survey. Edinburgh: E & S Livingstone, Ltd., 1969. 7. Selvin S, Janerich DT: Four factors influencing birth weight. British J Preventive Social Med 25:12-16, 1971. 8. Shapiro S, Schlesinger ER, Nesbitt REL: Infant, Perinatal, Maternal and Childhood Mortality in the United States. Cambridge: Harvard University Press, 1968. 9. Erhardt CL, Abramson H, Pakter J, et al: An epidemiological approach to infant mortality. Arch Environ Health 20: 743-757, 1970.
10. Kessner DM, Singer J, Kalk CE, et al: Infant Death: An Analysis by Maternal Risk and Health Care. Contrasts in Health Status, Vol. 1. Pub. No. OSBN 0-309-02119-7, Institute of Medicine, National Academy of Sciences, Washington, DC, 1973. 11. Gendell M, Hellegers E: The influence of the changes in maternal age, birth order and color on the changing perinatal mortality, Baltimore, MD, 1961-66. Health Services Reports 88:733742, 1973. 12. Dott AB, Fort AT: The effect of maternal demographic factors on infant mortality rates. Summary of the findings of the Louisiana Infant Mortality Study. Part I. Am J Obstet Gynecol 123:847-853, 1975. 13. Morris NM, Udry JR, Chase CL: Shifting age-parity distribution of births and the decrease in infant mortality. Am J Public Health 65:359-361, 1975. 14. Brazie JV, Pratt MW, Hexter AC, et al: Selected Natality Characteristics for Single Live Births, United States, 1974. DHEW Pub. No. (HSA) 79-5744, 1979. 15. Yerushalmy J, Van den Berg BJ, Erhardt CL, et al: Birthweight and gestation as indices of immaturity. Am J Dis Childr 109:4357, 1965. 16. Lubchenko LO: The High Risk Infant. Philadelphia: W.B. Saunders, 1976. 17. Williams RL: Intrauterine growth curves: Intra- and international comparisons with different ethnic groups in California. Preventive Medicine 4: 163-172, 1975. 18. Fleiss JL: Statistical Methods for Rates and Proportions. New York: Wiley, 1973. 19. Cox DR: The Analysis of Binary Data. London: Methuen, 1970. 20. Day NE, Kerridge DF: A general maximum likelihood discriminant. Biometrics 23: 313-323, 1967. 21. Halperin M, Blackwelder WC, Verter JI: Estimation of the multivariate logistic risk function: a comparison of the discriminant function and maximum likelihood approaches. J Chron Dis 24:125-158, 1971. 22. Mantel N, Haenszel W: Statistical aspects of the analysis of data from retrospective studies of disease. J National Cancer Institute 22:719-748, 1959. 23. Mantel N: Chi-square tests with one degree of freedom: extensions of the Mantel-Haenszel procedure. J Am Statist Assoc 58:690-700, 1963.
ACKNOWLEDGMENTS
Drs. Helen M. Wallace and Hyman Goldstein participated in the planning of this study. Mr. Naresh Sayal and Mr. Rao Yelamanchili were responsible for computer programming and tabulations. Mr. Eric Jaeger, Ms. Debra Kurita, Ms. Carol Royster and Mr. Kent Thomas compiled tables and did many calculations. Ms. Genevieve Coulson typed tables and manuscript. This work was supported in part by Grants No. MC-R-60208-05 and No. MC-R-1 10387-02 of the Office of Clinical Services, Bureau of Community Health Services, U.S. Department of Health, Education, and Welfare, Rockville, MD.
ATPM 1979 Annual Meeting Luncheon Sunday, November 4, 1979 New York Hilton Hotel Association of Teachers of Preventive Medicine. (For Department Chairpersons and Liaison Representatives only) Purchase tickets in advance by sending $15.50 to William M. Marine, MD, ATPM, University of Colorado Medical Center, No. C-245, 4200 E. 9th Avenue, Denver, CO 80262. Deadline date for ticket sales is October 15. Unmailed tickets will be held at the door. Make checks payable to ATPM. For information call 303/394-7502. AJPH September 1979, Vol. 69, No. 9
893
Abstract: Descriptive analyses of birthweight for single live births in the United States during 1974, using birth certificate information, show that several factors are associated with a high incidence of low birthweight babies. Multivariate analyses were performed to determine correlates of low birthweight. When other factors are held constant, race not white, previous reproductive loss, short interpregnancy interval, outof-wedlock birth, no prenatal care, and maternal age
under 18 years or over 35 years each increase the risk of having an infant of low birthweight. High birth order and maternal education under 12 years of schooling were not consistent risk factors for low birthweight. Odds ratios are presented and the method for combining these, to estimate the risk for an individual mother having a low-birthweight infant, is illustrated. (Am J Public Health 69:887-893, 1979.)
Recent studies of pregnancy outcome have shown a progressive decline of perinatal and infant mortality in all risk categories, including mortality for small preterm infants.' The decline has come at a time of change in many aspects of maternal and infant care: associated factors include improved perinatal services available to both mother and baby, improved care or prevention of specific diseases, and an increased availability of means for spacing or preventing pregnancies. In our study of birth certificate information to evaluate specific factors which increase the risk of a poor pregnancy outcome, we have followed the suggestion of Lewis, et al.,2 and use birthweight distribution as an indicator of pregnancy outcome. Although infant mortality rates remain the primary indicator, low birthweight is also an indicator of considerable interest. Low birthweight is associated with increased perinatal and infant mortality and morbidity, including adverse sequelae such as mental retardation and learning disabilities. Many previous studies have shown that birthweight, in-
fant mortality, and infant morbidity vary with such strongly interacting factors as maternal age, social class, previous reproductive history, and prenatal care.3-'0 Most studies, however, have not attempted to determine the specific effect of single factors. Gendell and Hellegers"I in a linked-records study were able to attribute one-fourth of the drop in Baltimore perinatal mortality over a five-year period to an increase in lower-order births and younger mothers. Kessner, et al. 10 attempted to isolate the effects of health care on infant mortality by analyzing separate groups divided by social and medical risk of high mortality. They concluded that within any risk group, mortality varied inversely with the amount of care received. Dott and Fort,'2 also using a linked-records study, held socioeconomic status and birthweight constant and showed that infant mortality varied inversely with the number of prenatal visits. Finally, Morris, Udry and Chase'3 used an adjusted-rate method to attribute 27 per cent of the fall of infant mortality from 1965 to 1972 to changes in age and parity of the mothers. The study from which the present report has been drawn'4 comprised an examination of United States birth registration data for single live births in 1974. For each race it related the proportion of births which were low birthweight (less than 2501 grams) to risk factors ascertainable from birth certificates. We summarize here the principal findings of that study and report the results of multivariate analyses which examine the risk of having a low birthweight baby by race, maternal age, wedlock status, prenatal care, and maternal education for primigravida, and by these factors and birth order, reproductive history, and interpregnancy interval for multigravida.
Address reprint requests to Joseph V Brazie, MD, MPH, Maternal and Child Health, Annex 4, Room 500, California Department of Health Services, 2151 Berkeley Way, Berkeley, CA 94704. Dr. Eisner is with the San Francisco Department of Public Health; Ms. Pratt is with the Information Sciences Research Institute, Silver Spring, MD; and Dr. Hexter is with the University of California School of Public Health, Berkeley. This paper, submitted to the Journal February 12, 1979, was revised and accepted for publication May 17, 1979. Editor's Note: See also editorials pages 851, 852.
A-JPH September 1979, Vol. 69, No. 9
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EISNER, ET AL
Methods The study universe consisted of all single live births resident and recorded in the 50 United States and the District of Columbia during 1974. The original data file, obtained from the National Center for Health Statistics, included all records from 16 states and a 50 per cent sample from the remainder. Thus, the basic file consisted of 1,989,128 records representing 3,101,117 births. Only 30 states reported all variables, and not all of the certificates for these states provided all information requested. Therefore, the multivariate analyses were based on a subset of the basic file consisting of 505,243 records with complete information on race, age of mother, onset of prenatal care, total birth order, maternal education, wedlock status and, for multigravida, interpregnancy interval and reproductive history.* Analyses were done by the Maternal and Child Health Studies Project of the Information Sciences Research Institute, Silver Spring, MD, with further analyses by the Collaborative Research Project of the Maternal and Child Health Program, School of Public Health, University of California, Berkeley. The ideal dependent variable for the multivariate analyses would be a combination of birthweight and gestational age.'5 However, the analyses for the descriptive portion of our study raised serious questions about the reliability of gestational age reporting. For example, 22 per cent of the births with reported gestational age less than 26 weeks weighed more than 2500 grams. Numerous studies, e.g., Lubchenko,'6 have shown that no births with true gestational age less than 26 weeks weigh more than 2500 grams. It appeared to us that the majority of the errors were in gestational age rather than birthweight and were concentrated in the group of greatest interest to us.** Preliminary analyses, using a combination of birthweight and gestational age as dependent variable, gave essentially the same results as those using birthweight alone. We concluded that, for our analyses, a dependent variable combining gestational age and birthweight gave no advantage over one using birthweight alone. *Comparison of marginal tabulations for the basic file with the file used for multivariate analyses showed small but statistically significant differences. The multivariate file had more births weighing less than 2501 grams, fewer with maternal age under 18 years, maternal education under 12 years of schooling, no prenatal visits or birth out-of-wedlock, and more with previous losses or interpregnancy interval less than six months. While we do not believe these affect our conclusions, the possibility that births with complete information are not representative of the universe of all births cannot be excluded. **The National Center for Health Statistics calculates gestational age from date of birth and date of last menstrual period (LMP). This procedure is much more reliable than the previous method of asking directly the number of completed weeks of gestation. However, Williams17 has shown that this procedure tends to misclassify certain births, especially those with very short interpregnancy intervals where LMP may refer to the previous birth. The number misclassified is small but enough to bias seriously the lower gestational age groups. 888
We defined two birthweight groups as dependent variables for the multivariate analyses: 1. birthweight less than 1501 grams, all gestations 2. birthweight less than 2501 grams, all gestations These two birthweight groups are not independent because all births less than 1501 grams are also included in births less than 2501 grams. However, only about 15 per cent of the births under 2501 grams are also under 1501 grams, and analyses of the two groups showed substantial differences. Our preliminary studies treated race as an independent variable which could take on different values in a single analysis. However, Black mothers were more likely than White mothers to have low birthweight babies at every level of every factor studied. For this reason we made separate analyses of each racial group (White, Black and Other) for primigravida and for multigravida. Regression-type analyses with more than 500,000 observations were beyond our resources; therefore we grouped the observations. The groupings used for the independent variables were: 1. Maternal Age: Under 18 years, 18-34 years, 35 years and over; 2. Wedlock Status: In-wedlock, Out-of-wedlock; 3. Prenatal Care: First trimester, Second trimester, Third trimester, No care (in some analyses these were grouped into "Some care" and "No care"); 4. Maternal Education: Under 12 years of schooling, 12 years of schooling and over; 5. Birth Order (for multigravida): Less than six, six and over; 6. Reproductive History (for multigravida): No previous reproductive losses, one or more previous losses; 7. Interpregnancy Interval (for multigravida): Less than six months, six months and over; These categories provided 1,296 possible combinations of independent variables (144 for primigravida, 1,152 for multigravida), of which 390 contained no births. Thus, we had 906 "data points" for the multivariate analyses. Initial screening showed only minor differences in per cent low birthweight for the three trimesters of onset of prenatal care, so these were grouped into "Some care" vs. "No care" for reporting the multivariate comparisons (Table 1). We used two methods to estimate the relative importance of each independent variable: multiple logistic regression (MLR), and the Mantel-Haenszel (M-H) procedure. MLR'8. '9 is a linear regression which uses the logarithm of the relative proportion of cases to non-cases as the dependent variable. It is closely related to discriminant analysis, and may be regarded as an extension of discriminant analysis to a broader class of models.20 21 The regression coefficient for each independent variable provides an estimate of the "odds ratio," the ratio between the chance an event will occur when a given factor is present to the chance it will occur when the factor is absent. For example, an odds ratio of 4.95 for the risk factor "No prenatal care" with respect to birthweight less than 2501 grams can be interpreted to mean that if all other factors are the same a woman who has not received prenatal care is about five times as likely to give birth AJPH September 1979, Vol. 69, No. 9
RISK OF LOW BIRTHWEIGHT
to an infant weighing less than 2501 grams as is a woman who has received some prenatal care. Similarly, an odds ratio less than 1.0 means the risk is reduced, rather than increased, if the factor is present. The M-H procedure22 23 is a chi-square-based method for combining contingency tables to assess the significance of a particular classifying variable in the presence of identified nuisance variables. It provides a chi-square with one degree of freedom, expressing whether low birthweight appears significantly more (or less) often in cases with the factor than in cases without the factor being studied. It also provides an estimate of the odds ratio, which may differ from the estimate provided by MLR. Because of time and budget limitations we did not test all categories of variables by both procedures.
of-wedlock being 6.4 per cent for Whites, 47.5 per cent for Blacks and 14.4 per cent for Others. Substantially higher rates for low birthweight babies were observed for out-ofwedlock births, for all races, with the proportionately greatest increase for Whites (Table 1). The M-H procedure, which was done for primigravida, showed a significantly increased risk of low birthweight for White and Black primigravida born out-of-wedlock but not for Other primigravida. Odds ratios for Whites and Blacks ranged from 1.17 to 1.45. Odds ratios by MLR for primigravida were similar. Odds ratios by MLR for multigravida were substantially higher than for primigravida for White and Other out-of-wedlock births (1.54 to 3.41) but not for Blacks (Table 3). Prenatal Care
Results The distribution of births and of the per cent of births less than 2501 grams are shown for each independent variable in Table 1. In considering the results it should be borne in mind that the "Other" race group is quite heterogeneous and comprised only about 2.5 per cent of all births. Therefore, while results are presented for the "Other" race group, with statements concerning statistical significance, the results for this group should be considered cautiously. The effectiveness of the MLR can be summarized by multiple correlation coefficients. In general, the regressions "explained" more of the variance for the primigravida than for the multigravida, for the births less than 1501 grams than for the births less than 2501 grams, and for the Whites than for the Blacks or Others (Table 2). The results of the M-H procedures and of the MLR are shown in Table 3. Maternal Age Black mothers were more likely to be under 18 years of age than White or Other mothers. Thus, Black mothers had substantially more births in the age group with the highest proportion of low birthweight babies (Table 1). The MLR showed that for all three races primigravida over 35 years of age had substantially increased risk of low birthweight compared to mothers aged 18-34 years; odds ratios ranged from 2.00 to 8.86. Multigravida over 35 years of age also had increased risk but not so great as for primigravida. The M-H procedure was not done for this variable (Table
3).
Maternal age under 18 years was a substantial risk factor for multigravida, with odds ratios ranging from 1.34 to 6.72. Young maternal age was not an important factor for primigravida when other factors were held constant; odds ratios ranged from not significantly different from 1.0 to a maximum of 2.74. Wedlock Status A substantial difference for proportion of births by wedlock status is evident between races, with the per cent outAJPH September 1979, Vol. 69, No. 9
Black mothers tended to have their first visit for prenatal medical care somewhat later than did White or Other mothers. However, for all three race groups the median time, for those with time of first visit reported, was in the first trimester. The time of first prenatal visit made very little difference in per cent of births which were low birthweight, provided the mother had at least one prenatal visit. However, no prenatal medical care was a very significant risk factor for the 1.4 per cent of mothers who lacked it. No prenatal care was the greatest risk factor for low birthweight in our study. The M-H procedure showed highly significant increased risks of birthweights below 1501 grams and below 2501 grams for primigravida, and for mothers with no prenatal care in all three race groups; odds ratios ranged from 2.78 to 15.65. Odds ratios from MLR were similar to those from M-H, for primigravida. Odds ratios by MLR for multigravida with no prenatal medical care were statistically significant, ranging from 2.09 to 5.88. In every case, the odds ratios for multigravida were substantially lower than for the corresponding group of primigravida. The M-H procedure was not done for multigravida, for this variable (Table 3).
Maternal Education Most mothers had achieved 12 years of education (high school graduation) or more: 74.5 per cent for Whites, 53.6 per cent for Blacks, and 67.1 per cent for Others. Achievement of at least high school graduation was associated with improved outcome in terms of low birthweight rates (Table 1).
The multivariate analyses gave paradoxical results for this factor. The M-H procedure showed strong associations between maternal education under 12 years of schooling and birthweights below both 1501 grams and 2501 grams for White primigravida and for birthweights below 2501 grams for Black primigravida but not for Other primigravida. The same pattern occurred for multigravida. The MLR, however, gave quite different results. Of the six odds ratios calculated for primigravida, only that for White primigravida was statistically significant. For multigravida four of the six odds ratios were statistically significant, and all four were in the opposite direction to that expected (Table 3). 889
EISNER, ET AL TABLE 1 -Number of Births and Per Cent Le than 2501 Grams for Selected Variables by Race for Single Live Births, United States, 1974 Number of Births
Variable
Total
TOTAL 3,101,117 Maternal Age (years) Less than 15 12,436 15-17 231,691 18-19 356,461 20-24 1,089,049 25-29 903,573 30-34 363,872 35-39 115,080 40-44 27,260 45 and over 1,695 Wedlock Status In-wedlock 1,812,642 Out-of-wedlock 280,682 Not stated 10,570 Not reported' 997,223 Prenatal Care First trimester 1,852,701 Second trimester 557,092 Third trimester 123,935 No prenatal care 35,871 Not stated 109,994 Not reported' 421,524 Matemal Education (years) 0-8 134,794 9-11 508,463 12 994,045 13-15 325,635 16 and over 239,482 Not stated 59,172 Not reported' 839,526 Birth Order First 1,220,082 Second 926,821 Third 451,258 Fourth 214,679 Fifth 105,515 Sixth 55,559 Seventh 31,698 Eighth and over 48,016 Not stated 47,489 Reproductive History No previous losses 1,311,558 One or more previous losses 405,902 No previous pregnancy 1,146,894 Not stated 236,763 Interpregnancy Interval Less than 6 months 116,514 6-11 months 133,685 12-23 months 263,377 24 months and over 605,387 No previous pregnancy 964,762 Not stated 465,106 Not reported' 552,286
Per Cent Less than 2501 Grams
White
Black
Other
Total
White
Black
Other
2,529,411
495,950
75,756
6.5
5.4
12.0
6.5
5,010 150,777 264,655 896,737 786,497 309,583 93,248 21,591 1,313
7,241 76,987 85,009 169,925 92,467 41,569 17,610 4,802
185
3,927 6,797 22,387 24,609 12,720 4,222
15.3 10.5 8.4 6.2 5.2 5.8 7.1 8.4 9.6
12.2 8.5 6.9 5.2 4.6 5.2 6.3 7.5 9.4
17.5 14.5 13.3 11.4 10.1 10.6 11.5 12.2 10.6
9.2 8.3 8.0 6.4 5.8 6.2 7.8 10.0 7.1
1,590,022 108,050 9,109 822,230
183,240 166,031 1,285 145,394
29,599
5.8 11.9 6.9 6.3
5.2 9.1 6.1 5.4
10.4 13.9 12.1 11.9
6.4 8.4 9.1 6.3
1,592,762 400,029
217,871 140,049 34,324 12,412 25,404 65,890
42,068 17,014 5,214 1,389 1,697 8,374
5.7 7.7 7.8 18.9 8.7 6.7
4.9 6.2 6.4 15.3 6.8 5.7
11.0 12.1 11.5 25.7 14.7 11.8
6.2 6.8 5.7 14.5 8.2 6.5
29,810 143,084
5,275 11,132
147,812 36,187 15,430 13,639 109,988
16,667 7,134 9,640 1,238 24,670
9.3 9.4 5.9 5.0 4.2 7.8 6.2
8.0 7.7 5.0 4.3 3.8 6.0 5.4
14.0 13.6 11.0 10.1 8.7 13.7 11.6
7.2 7.4 6.5 5.9 6.0 6.5 6.4
775,851 369,776 171,143 81,853 41,643 22,963 31,999 36,588
193,015 128,104 70,360 38,466 21,012 12,461 7,886 14,569 10,077
29,472 22,866 11,122 5,070 2,650 1,455 824
6.9 5.8 6.1 6.8 7.3 7.9 8.0 8.4 6.8
5.9 4.8 5.0 5.6 6.3 6.8 6.7 7.1 5.3
12.5 11.7 11.6 11.9 11.5 11.6 11.5 11.4 12.3
6.9 6.2 6.0 6.5 6.5 7.5 8.1 6.6 6.8
1,083,255
195,368
32,935
5.4
4.5
10.4
5.5
324,500
70,707
10,695
9.0
7.6
15.2
8.7
940,840 180,816
177,671 52,204
28,383
3,743
6.9 6.7
5.9 5.1
12.5 12.1
7.0 6.0
92,979 108,287 223,994
19,341 20,832 32,102
4,194 4,566 7,281
7.8 6.3 5.1
6.4 5.2 4.3
14.5 12.1 10.7
8.1 6.7 5.7
502,851
88,961
13,575
5.2
4.4
9.7
4.7
793,079 355,287 452,934
145,003 97,321 92,390
26,680 12,498 6,962
5.8 6.3 5.8
12.5 13.3
6.9 7.6 6.9
84,397 22,070 82,893 347,260
99,709 354,247 829,566 282,314 214,412 44,295 704,868
997,595
340
867 42
39,380 6,601 176
849
1,448
6.8
7.8 6.8
11.9
'This item not reported on birth certificates in a number of states.
890
AJPH September 1979, Vol. 69, No. 9
RISK OF LOW BIRTHWEIGHT
TABLE 2-Multiple Correlations (R2) for Multiple Logistic Regressions Birthweight Group Race and Gravidity
Primagravida White Black Other Multigravida White Black Other
Less than 1501 grams
Less than 2501 grams
.863 .725 .663
.770 .512 .679
.600 .540 .558
.422 .377 .322
Birth Order
Primigravida comprised about 40 per cent of the births in all racial groups. The numbers decreased rapidly with increasing birth order, more rapidly for Whites than for the other two race groups; 7.2 per cent of the Black births were birth order six or higher, compared to 3.9 per cent for Whites and 5.0 per cent for Others. For Whites and Others the lowest per cent low birthweight was observed for birth orders two to four (two to five for Others). For Blacks the low birthweight per cent was 12.5 per cent for first births, and lower for the higher order births (Table 1). The effect of high birth order (gravida six and over) was examined using MLR. When other variables were taken into account the effect of high birth order was small (Table 3). Reproductive History Percentages of multigravida with one or more previous losses prior to this delivery were 23.1 per cent for Whites, 26.6 per cent for Blacks, and 24.5 per cent for Others. The association of reproductive history with the birthweight outcome of the current pregnancy is striking. Each race showed about a 50 per cent increase in per cent low birthweight for one or more previous losses, compared to no previous losses. The low birthweight rates for primigravida were intermediate (Table 1). The M-H procedure indicated a statistically highly significant increased risk of low birthweight for mothers with one or more previous losses for all three race groups, for both birthweight categories; the odds ratios ranged from 1.56 to 2.69. Similar results were obtained by the MLR, with odds ratios of 1.34 to 2.22 (Table 3).
Interpregnancy Interval The per cent of multigravida mothers with interpregnancy interval less than six months was 10.0 per cent for Whites, 12.0 per cent for Blacks, and 14.2 per cent for Others. A short interpregnancy interval was associated with an increased per cent of low birthweight babies. The lowest proportions of low birthweight babies occurred to mothers who waited at least 12 months after the previous delivery before again becoming pregnant (Table 1). The M-H procedure indicated an increased risk of low birthweight for both birthweight categories to White and AJPH September 1979, Vol. 69, No. 9
Black multigravida with interpregnancy intervals of less than six months; odds ratios ranged from 1.12 to 1.55. The odds ratio for Other multigravida was substantial for births less than 1501 grams (1.81) but was not significantly different from 1.0 for births less than 2501 grams. MLR indicated substantially increased risk of low birthweight for all three race groups and both birthweight categories; odds ratios by MLR ranged from 1.31 to 2.69 and in every case were higher than the odds ratios by the M-H procedure (Table 3).
Discussion and Conclusions This study supports the findings of previous studies which associate a poor outcome of pregnancy with certain risk factors. It also permits certain conclusions for program planning. Both the regression (MLR) analyses and the MantelHaenszel (M-H) procedure show that a previous pregnancy loss, an interpregnancy interval less than six months, a birth out-of-wedlock, and a pregnancy without prenatal medical care all increase the risk of having a low birthweight baby, independent of each other and of the other variables studied. The magnitude of the increased risk varies with gravidity and race. In general the greatest risk comes from pregnancies with no prenatal medical care, and the risks from the other factors, while appreciable, are considerably smaller. The regression analyses of maternal age show a higher risk for multigravida under 18 years of age than for primigravida in this age group, and a higher risk for primigravida over 35 years of age than for similarly aged multigravida. These findings, and the inconsistently higher risks for the sixth and higher birth orders, conform with conventional wisdom. The two methods of analysis showed contradictory results for the effect of maternal education. This occurred because the two forms of multivariate analysis measure different things. The M-H procedure, like most chi-squarebased tests, is primarily reporting the largest groups. MLR, which looks only at the proportions, like most regression procedures gives greatest weight to observations furthest from the mean and therefore is primarily reporting consistency. From a public health standpoint the results of M-H, which showed low maternal education to be a risk factor for Whites for both birthweight groups, and for Blacks for birthweights below 2501 grams, would appear to be more meaningful. The regression results indicating a contrary effect suggest that social and environmental effects are operating which were not identified by birth certificate information and are not randomly distributed over the categories used. However, the odds ratios derived from maternal education are small. If this factor has an independent effect, the effect is of little importance. This study indicates that lack of prenatal care is an important risk indicator for low birthweight. Lack of prenatal care, however, not only indicates the possibility that needed medical intervention did not occur, it also may indicate nonconventional health behavior, including maternal behaviors directly detrimental to the infant. Also, an unknown proportion of births with "no care" represent premature deliveries 891
EISNER, ET AL TABLE 3-Results of Mantel-Haenszel Procedures and Multiple
Logistic Regressions Odds Ratio
Chi-square' from Mantel-Haenszel Risk Factor, Race, and Gravidity
Birth less than 1501 grams
Maternal age less than 18 years White primigravida Black primigravida Other primigravida White multigravida Black multigravida Other multigravida Maternal age 35 years and over White primigravida Black primigravida Other primigravida White multigravida Black multigravida Other multigravida Birth out-of-wedlock White primigravida 29.33** Black primigravida 5.88* Other primigravida 0.33 White multigravida Black multigravida Other multigravida No prenatal care White primigravida 710.27** Black primigravida 361.36** Other primigravida 42.53**
White multigravida
Black multigravida Other multigravida Maternal education less than 12 years White primigravida 7.03** Black primigravida 2.82 Other primigravida 0.16 White multigravida 28.41" Black multigravida 0.72 Other multigravida 0.18 Birth order six and over White multigravida Black muftigravida Other muftigravida One or more previous pregnancy losses White multigravida 226.04** Black muftigravida 125.21** Other multigravida 15.32** Interpregnancy interval less than six months White multigravida 21.09** Black multigravida 19.14** Other multigravida 4.49*
Birth less than 2501 grams
63.18** 15.78** 0.04
from Mantel-Haenszel2 Birth less than 1501 grams
1.45** 1.23* 1.38
Birth less than 2501 grams
1.29** 1.1 7** 1.04
from regressions Birth less than 1501 grams
2.71
4.10** 5.77** 8.86** 2.45** 2.19** 2.63**
2.00" 3.56" 1.27" 1.17 1.70"
1.41 *
1.12
1.58*
1.52' 1.09 1.76" 1.28" 1.54" 4.95*' 2.83" 5.91" 2.92" 2.09" 2.75"
1.53 1.24*
2.25" 11.53** 6.19** 15.65**
4.92** 2.78** 4.80**
14.97" 5.05" 14.29" 5.88" 3.45" 4.04"
117.64** 23.52** 0.01 491.22** 12.89** 0.44
1.20** 1.17 1.27 1.38** 0.92 0.86
1.36** 1.24** 1.02 1.71 ** 1.13** 0.93
1.21 0.96
1.79** 1.16 2.74** 4.38** 2.93** 6.72**
3.41" 451.67** 179.08** 20.83**
Birth less than 2501 grams
1.07 1.12 0.95
0.72" 0.52"
1.65' 1.63" 1.34"
2.11
1.32* 0.97 0.88 1.01
0.84*
0.73*
0.82
1.58"
0.98 0.85 1.38*
0.97
2.21" 484.04** 175.62** 14.98**
2.33** 2.61 ** 2.69**
1.63** 1.66** 1.56**
1.63" 2.22" 1.52"
1.54" 1.71"
15.91 **
1.38** 1.55** 1.81*
1.12** 1.28** 1.29
2.33" 2.69" 2.12"
1.31" 1.60" 1.57"
31.44** 3.37
1.34*
'All chi-squares have one degree of freedom. 2Significance levels for odds ratios from Mantel-Haenszel are by convention taken to be the same as for the associated chi-square.
*.01 < p < .05 **p < .01
to women who would have obtained care in the third trimester had the pregnancy not ended. This study showed that lack of prenatal care is an indicator of risk but does not allow conclusions about the effect of providing earlier care. In the same way, measures to increase interpregnancy intervals may contribute to substantial reductions in risk but our study 892
does not provide direct proof of this. The study shows that women with a previous pregnancy loss remain at high risk independent of their age, race, education, interpregnancy interval, gravidity, or wedlock status. While primigravid women pregnant out-of-wedlock are not necessarily at high risk, a woman who has a second (or higher) pregnancy out-of-wedAJPH September 1979, Vol. 69, No. 9
RISK OF LOW BIRTHWEIGHT
lock has a markedly increased risk of a low birthweight delivery. The expression of risks as odds ratios allows an estimate of individual risk by multiplying together the odds ratios for each risk factor borne by the individual. For example, if we assume risk factors are independent and the MLR odds ratios from this study apply, the risk of having an infant under 2501 grams for an unmarried White multigravida over 35 years of age, with a previous loss, can be estimated as 1.76 x 1.27 x 1.54 = 3.44 times the risk for a White multigravida who is married, 18-34 years of age, with no previous losses, all other factors being equal. In terms of total contribution to the number of infants with low birthweight in our study, 401 White unmarried multigravida over 35 years of age, with a previous loss, had 40 low birthweight infants compared to the 6,505 low birthweight infants delivered to the 164,805 White married multigravida 18-34 years of age, with no previous loss. Thus, this high risk group actually made a very small contribution to the overall problem, even though individually they were at substantially higher risk.
REFERENCES
1. Eisner V, Pratt MW, Hexter A, et al: Improvement in infant and perinatal mortality in the United States, 1%5-1973: I. Priorities for intervention. Am J Public Health 68:359-364, 1978. 2. Lewis R, Charles M, Patwary KM: Relationships between birth weight and selected social, environmental and medical care factors. Am J Public Health 63:973-981, 1973. 3. Shapiro S, Unger J: Weight at Birth and Its Effect on Survival of the Newborn in the United States, Early 1950. Vital StatisticsSpecial Reports, Vol. 39 No. 1. U.S. National Office of Vital Statistics, Public Health Service, Washington, DC, 1954. 4. Shapiro S, Unger J: Relation of Weight at Birth to Cause of Death and Age at Death in the Neonatal Period: United States, Early 1950. Vital Statistics-Special Reports, Vol. 39 No. 6. U.S. National Office of Vital Statistics, Public Health Service, Washington, DC, 1956. 5. Butler NR, Bonham DG: Perinatal Mortality: The First Report of the British Perinatal Mortality Survey. Edinburgh: E & S Livingstone, Ltd., 1%3. 6. Butler NR, Alberman ED, (Eds.): Perinatal Problems: The Second Report of the British Perinatal Mortality Survey. Edinburgh: E & S Livingstone, Ltd., 1969. 7. Selvin S, Janerich DT: Four factors influencing birth weight. British J Preventive Social Med 25:12-16, 1971. 8. Shapiro S, Schlesinger ER, Nesbitt REL: Infant, Perinatal, Maternal and Childhood Mortality in the United States. Cambridge: Harvard University Press, 1968. 9. Erhardt CL, Abramson H, Pakter J, et al: An epidemiological approach to infant mortality. Arch Environ Health 20: 743-757, 1970.
10. Kessner DM, Singer J, Kalk CE, et al: Infant Death: An Analysis by Maternal Risk and Health Care. Contrasts in Health Status, Vol. 1. Pub. No. OSBN 0-309-02119-7, Institute of Medicine, National Academy of Sciences, Washington, DC, 1973. 11. Gendell M, Hellegers E: The influence of the changes in maternal age, birth order and color on the changing perinatal mortality, Baltimore, MD, 1961-66. Health Services Reports 88:733742, 1973. 12. Dott AB, Fort AT: The effect of maternal demographic factors on infant mortality rates. Summary of the findings of the Louisiana Infant Mortality Study. Part I. Am J Obstet Gynecol 123:847-853, 1975. 13. Morris NM, Udry JR, Chase CL: Shifting age-parity distribution of births and the decrease in infant mortality. Am J Public Health 65:359-361, 1975. 14. Brazie JV, Pratt MW, Hexter AC, et al: Selected Natality Characteristics for Single Live Births, United States, 1974. DHEW Pub. No. (HSA) 79-5744, 1979. 15. Yerushalmy J, Van den Berg BJ, Erhardt CL, et al: Birthweight and gestation as indices of immaturity. Am J Dis Childr 109:4357, 1965. 16. Lubchenko LO: The High Risk Infant. Philadelphia: W.B. Saunders, 1976. 17. Williams RL: Intrauterine growth curves: Intra- and international comparisons with different ethnic groups in California. Preventive Medicine 4: 163-172, 1975. 18. Fleiss JL: Statistical Methods for Rates and Proportions. New York: Wiley, 1973. 19. Cox DR: The Analysis of Binary Data. London: Methuen, 1970. 20. Day NE, Kerridge DF: A general maximum likelihood discriminant. Biometrics 23: 313-323, 1967. 21. Halperin M, Blackwelder WC, Verter JI: Estimation of the multivariate logistic risk function: a comparison of the discriminant function and maximum likelihood approaches. J Chron Dis 24:125-158, 1971. 22. Mantel N, Haenszel W: Statistical aspects of the analysis of data from retrospective studies of disease. J National Cancer Institute 22:719-748, 1959. 23. Mantel N: Chi-square tests with one degree of freedom: extensions of the Mantel-Haenszel procedure. J Am Statist Assoc 58:690-700, 1963.
ACKNOWLEDGMENTS
Drs. Helen M. Wallace and Hyman Goldstein participated in the planning of this study. Mr. Naresh Sayal and Mr. Rao Yelamanchili were responsible for computer programming and tabulations. Mr. Eric Jaeger, Ms. Debra Kurita, Ms. Carol Royster and Mr. Kent Thomas compiled tables and did many calculations. Ms. Genevieve Coulson typed tables and manuscript. This work was supported in part by Grants No. MC-R-60208-05 and No. MC-R-1 10387-02 of the Office of Clinical Services, Bureau of Community Health Services, U.S. Department of Health, Education, and Welfare, Rockville, MD.
ATPM 1979 Annual Meeting Luncheon Sunday, November 4, 1979 New York Hilton Hotel Association of Teachers of Preventive Medicine. (For Department Chairpersons and Liaison Representatives only) Purchase tickets in advance by sending $15.50 to William M. Marine, MD, ATPM, University of Colorado Medical Center, No. C-245, 4200 E. 9th Avenue, Denver, CO 80262. Deadline date for ticket sales is October 15. Unmailed tickets will be held at the door. Make checks payable to ATPM. For information call 303/394-7502. AJPH September 1979, Vol. 69, No. 9
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