We examined the association between survival of infants with severe congenital heart defects (CHDs) and community-level indicators of socioeconomic status.
We identified infants born to residents of Arizona, New Jersey, New York, and Texas between 1999 and 2007 with selected CHDs from 4 population-based, statewide birth defect surveillance programs. We linked data to the 2000 US Census to obtain 11 census tract–level socioeconomic indicators. We estimated survival probabilities and hazard ratios adjusted for individual characteristics.
We observed differences in infant survival for 8 community socioeconomic indicators (
The increased mortality risk among infants with CHDs living in socioeconomically deprived communities might indicate barriers to quality and timely care at which public health interventions might be targeted.
Advances in medical and surgical care for individuals born with congenital heart defects (CHDs) has improved survival in recent years, yet despite this progress, mortality due to CHDs remains a significant public health issue.
Community-level factors related to socioeconomic conditions have been associated with decreased access to pediatric subspecialty care and early mortality of infants with low birth weight,
We used population-based data from 4 state-based birth defect surveillance programs (Arizona, New York, New Jersey, and Texas) to conduct a retrospective cohort study. The study population included live-born infants delivered from 1999 to 2007 with a diagnosis of 1 of the following 7 CHDs: common truncus arteriosus, transposition of the great vessels, tetralogy of Fallot, atrioventricular septal defect, aortic valve stenosis, hypoplastic left heart syndrome, and coarctation of the aorta. We selected these defects for inclusion in the analysis because of the high reliability with which they are ascertained by public health birth defect surveillance programs and because of the relatively high mortality associated with each defect. We classified infants as having one of the included CHDs by using a modified British Pediatrics Association (BPA) coding system
We matched infants with CHDs identified by state surveillance programs to state-specific linked birth-infant death files to determine vital status and to retrieve sociodemographic variables (state, county, and census tract number of the maternal residence at birth, maternal race, maternal nativity, maternal education, and maternal age) and clinical variables (infant’s birth weight, parity, and infant’s sex). Census tracts are small-area groupings (approximately 4000 residents) consisting of relatively homogeneous population characteristics. We obtained census tract information by linking individual-level data to the 2000 US Census by the census tract number of the maternal residence, and then extracted the following 11 census tract–level socioeconomic variables, selected through a literature review
Using the Kaplan-Meier product-limit method, we estimated survival probabilities for infancy (0–364 days) and for the neonatal (< 28 days) and postneonatal (28–364 days) periods.
To examine the possibility that each community SES indicator was an observable measure of a common underlying risk factor, we estimated survival probabilities for an index score, which was based on the cumulative number of indicators for which the infant’s census tract was in the lowest, or most disadvantaged, decile. On the basis of the distribution of the number of indicators in the most disadvantaged decile, we created levels 1 through 4 of the variable, consisting of infants who had 0, 1, 2 to 4, or 5 or more of the 11 socioeconomic indicators in the lowest decile, respectively.
For the census tract indicators that showed an association with survival and for which there was a corresponding individual-level variable (i.e., education, race/ethnicity), we stratified survival estimates by the individual-level variable to examine whether the magnitude of risk associated with the community-level variable was consistent across individual-level risk factors.
We used Cox proportional hazard regression models to estimate the effect of census tract–level socioeconomic factors on mortality, controlling for individual-level variables that were statistically significant in the univariate analyses.
We also used crude proportional hazard regression models to assess the relationship between maternal race/ethnicity and survival. We then adjusted these models with community measures of SES to examine whether the estimates of racial/ethnic disparities in survival were attenuated. All proportional hazard regression models for the postneonatal period assumed survival through the first 27 days. We performed computations using SAS version 9.2 (SAS Institute, Cary, NC).
Overall, we identified 10 578 infants with at least 1 of the 7 selected CHDs from the 4 population-based birth defect surveillance programs. Of those, 9853 infants (93%) had census tract information that could be used for linkage with the 2000 US Census. Maternal and infant characteristics of the final cohort by participating state are provided in Table A and infant mortality by CHD type are provided in Figure A (both available as supplements to the online version of this article at
The overall infant survival was 80.3% (95% confidence interval [CI] = 79.5%, 81.1%) (
We stratified community indicators for race/ethnicity and education by the corresponding individual-level factor. Being born to a mother with less than a high school education was associated with poorer infant survival (
Individual-level covariates that were associated with survival were birth weight, infant’s sex, maternal age, maternal nativity, maternal education, parity, birth period, and state of residence. There were significant interactions between maternal age and 2 SES indicators: the proportion speaking a language other than English and the proportion foreign born; both were more strongly associated with infant mortality among older mothers. There were additional interactions between maternal race/ethnicity and 2 SES indicators: the proportion of the population in an operator or laborer occupation was more strongly associated with non-Hispanic White infants, and residential crowding was more strongly associated with Hispanic ethnicity. We also found an interaction between per capita income and infant birth weight. The greatest infant mortality risk was among Hispanic infants for residential crowding (adjusted hazard ratios [AHR] = 4.24; 95% CI = 1.56, 11.54;
To examine the increased mortality risk associated with extreme socioeconomic disadvantage, we modeled a composite socioeconomic variable, as described in Methods, to determine the increased risk of living in communities with more indicators of socioeconomic disadvantage, adjusting for individual-level factors. Compared with infants living in a census tract with none of the indicators in the lowest decile, infants living in a census tract with 1 indicator, 2 to 4 indicators, and 5 or more indicators in the lowest decile had 10%, 15%, and 25% increased mortality risk, respectively, after adjustment for individual-level factors (
The greatest observed racial disparity was in the postneonatal period, during which crude mortality risk was 86% higher for non-Hispanic Blacks and 57% higher for Hispanics compared with non-Hispanic Whites (data not shown). Statistical adjustment for individual-level factors reduced the excess postneonatal mortality risk to non-Hispanic Blacks by 29% and the excess mortality risk to Hispanics by 23%. Adjustment for only the census tract–level measures of socioeconomic factors that were associated with survival had no notable impact on the crude hazard ratio for non-Hispanic Blacks and Hispanics.
Socioeconomic disadvantage was adversely associated with the survival of infants born with CHDs. Survival varied by most but not all community indicators that we examined, and the statistical significance of the survival difference was stronger in the postneonatal period than in the neonatal period. The community factors most predictive of infant death were related to income, poverty, education, and occupation. Socioeconomic disadvantage related to these factors increased the infant mortality risk by up to 47%, and the associated mortality risk increased significantly in some subpopulations, such as those of Hispanic ethnicity. Among infants born to Hispanic mothers, those who lived in communities with high residential crowding had more than a fourfold increased infant mortality risk compared with those living in communities with the least residential crowding.
Because infants with severe CHDs require early and continued surgical and medical intervention, increased access to and use of specialized health care resources would be expected to improve the likelihood of survival; however, measuring access or barriers to care in population-based studies is challenging.
CHD subtypes can range in severity. Those included in this study are considered severe and were selected because of the reliability with which they are clinically diagnosed and accurately detected by birth defect surveillance programs. Because of the severe nature of these conditions, infants require surgical intervention and appropriate follow-up care, with additional surgeries frequently required. The complexity of continued care might explain why the impact of SES was greater on postneonatal survival. Higher levels of education might be needed to understand and process sophisticated medical information and to research and make selection decisions about the physicians and institutions that might provide higher-quality care. Family income is associated with gradients in both children’s health and access to health care, and it might influence the ability to seek out and use high-quality care, especially to overcome barriers presented by significant travel distances.
The finding that occupation influenced survival was interesting and potentially more difficult to explain. This variable might serve as a proxy for health literacy or family resources, although it was not highly correlated with the measures for income and education. It is more likely that occupation type is an indicator of the level of medical insurance coverage, which is often employer based. Unemployment was not associated with mortality; however, Medicaid eligibility might be higher in areas of high unemployment, with covered individuals having sufficient insurance coverage and access to care. Those more likely to be in operator or laborer occupations might have incomes too high to qualify for Medicaid but have no employer-provided coverage or have plan options that provide inadequate coverage. The impact of medical insurance on infant survival among individuals born with CHDs has not been well examined, although several hospital-based studies found that infants with public insurance had higher postoperative mortality than infants covered by private insurance.
The consistency with which we observed lower infant survival for the most disadvantaged decile across all the SES indicators raised the issue of whether each indicator was indicative of some common underlying factor associated with economic disadvantage. If that were the case, having multiple indicators in the lowest decile would not be expected to increase the risk above that associated with a single indicator; however, we found a decrease in infant survival with an increasing number of indicators that were in the most disadvantaged decile. This suggests that different community indicators of SES might be separable measures of differing underlying risks, and analyses limited to 1 community indicator might underestimate the impact of disadvantaged communities on health outcomes. Another unknown was whether the community indicators were simply reflecting an individual-level risk or if the community risk was independent of the individual risk. For the most direct test, we examined constructs for which both the individual and community variables were associated with survival. Low maternal education was associated with lower survival, but even infants born to mothers with higher education had significantly lower survival if they lived in a community with a high proportion of residents with a low level of education. In fact, the community effect associated with low education appeared to be larger than the individual-level effect. Similarly, even though non-Hispanic White infants had a survival advantage compared with Hispanic infants, non-Hispanic White infants in communities with a high proportion of Hispanic residents experienced lower survival than non-Hispanic White infants living in communities with a low proportion of Hispanic residents, which is in notable contrast to the “Hispanic paradox” observed in the general population.
The observed racial/ethnic disparities in mortality among infants with CHDs corroborate earlier findings that have not been well explained.
A strength of this study was that it combined data from several states to provide a larger study population size that was diverse racially and ethnically, regionally, and socioeconomically; however, a disproportionate number of infants were from urban communities. This large, diverse population reduced the level of random variability in the survival estimates and allowed evaluation of differences in these estimates across an array of factors that might influence survival of infants with CHDs. Birth defect surveillance programs use varying methodologies to identify and confirm cases of birth defects, and the programs from which these data were drawn are those that employ active or partially active case ascertainment, which provide more accurate clinical diagnoses and the most complete prevalence estimates.
A potential limitation of this study was the lack of information on pregnancies affected by a CHD that resulted in fetal deaths, especially those that were elective terminations. Socioeconomic and other barriers to prenatal care can result in differential prenatal diagnosis by race and SES.
Although birth defects are a leading cause of infant death, public health strategies to address the overall burden of infant deaths have largely been concentrated on other causes such as preterm delivery, sudden infant death syndrome, and injuries.
We thank Gang Liu, MS, for his invaluable contribution of obtaining the census tract data for all states. We also thank the entire staff of the New York State Congenital Malformations Registry within the New York State Department of Health; Mark Canfield, PhD, and the entire staff of the Texas Birth Defects Registry within the Texas Department of State Health Services; Timothy Flood, MD, and the entire staff of the Arizona Birth Defects Monitoring Program within the Arizona Department of Health Services; and Leslie Beres and the entire staff of the Special Child Health Services Registry within the New Jersey Department of Health. Without these agencies, these data could not have been obtained.
J. E. Kucik conceptualized and designed the study, carried out the analyses, drafted the initial manuscript, and approved the final manuscript as submitted. W. N. Nembhard, P. Donohue, O. Devine, C. S. Minkovitz, T. Burke, and Y. Wang contributed to the conceptualization and study design, reviewed and revised the manuscript, and approved the final manuscript as submitted.
This study received institutional review board approvals from the Centers for Disease Control and Prevention and the Texas Department of State Health Services.
| Neonatal | Postneonatal | Infant | ||||||||
|---|---|---|---|---|---|---|---|---|---|---|
| Census Tract Variable | No. of Infants Born | No. of Deaths | Survival, % |
| No. of | Survival, % |
| No. of | Survival, % |
|
| Total | 9853 | 1019 | 89.7 (89.0, 90.2) | 923 | 89.6 (88.9, 90.2) | 1942 | 80.3 (79.5, 81.1) | |||
| Per capita income | .02 | .003 | < .001 | |||||||
| Most disadvantaged decile | 1044 | 125 | 88.0 (85.9, 89.9) | 122 | 86.7 (84.4, 88.8) | 247 | 76.3 (73.6, 78.7) | |||
| Least disadvantaged decile | 633 | 52 | 91.8 (89.4, 937) | 42 | 92.8 (90.3, 94.6) | 94 | 85.2 (82.1, 87.7) | |||
| Poverty | .01 | .004 | < .001 | |||||||
| Most disadvantaged decile | 924 | 123 | 86.7 (84.3, 88.7) | 96 | 88.0 (85.36, 90.1) | 219 | 76.3 (73.4, 78.9) | |||
| Least disadvantaged decile | 734 | 69 | 90.6 (88.2, 92.5) | 50 | 92.5 (90.2, 94.2) | 119 | 83.8 (80.9, 86.3) | |||
| Education | .01 | <.001 | < .001 | |||||||
| Most disadvantaged decile | 1203 | 142 | 88.2 (86.2, 89.9) | 138 | 87.0 (84.8, 88.9) | 280 | 76.7 (74.2, 79.0) | |||
| Least disadvantaged decile | 765 | 63 | 91.8 (89.6, 93.5) | 50 | 92.9 (90.7, 94.6) | 113 | 85.2 (82.5, 87.6) | |||
| Operator/laborer occupation | .004 | .004 | < .001 | |||||||
| Most disadvantaged decile | 1420 | 161 | 88.7 (86.9, 90.2) | 143 | 88.6 (86.8, 90.3) | 304 | 78.6 (76.4, 80.6) | |||
| Least disadvantaged decile | 637 | 45 | 92.9 (90.7, 94.7) | 42 | 92.9 (90.5, 94.7) | 87 | 86.3 (83.4, 88.8) | |||
| Hispanic, % | .42 | < .001 | .003 | |||||||
| Most disadvantaged decile | 1393 | 169 | 87.9 (86.0, 89.5) | 167 | 86.4 (84.3, 88.2) | 336 | 75.9 (73.5, 78.0) | |||
| Least disadvantaged decile | 968 | 106 | 89.1 (86.9, 90.9) | 63 | 92.7 (90.7, 94.2) | 169 | 82.5 (80.0, 84.8) | |||
| Non-English speaking | .33 | .003 | .007 | |||||||
| Most disadvantaged decile | 1099 | 145 | 86.8 (84.7, 88.7) | 131 | 86.3 (83.9, 88.3) | 276 | 74.9 (72.2, 77.3) | |||
| Least disadvantaged decile | 1212 | 143 | 88.2 (86.3, 89.9) | 102 | 90.5 (88.5, 92.1) | 245 | 79.8 (77.4, 81.9) | |||
| Residential crowding | .31 | .004 | .006 | |||||||
| Most disadvantaged decile | 1129 | 126 | 88.8 (86.9, 90.5) | 130 | 87.0 (84.8, 89.0) | 256 | 77.3 (74.8, 79.7) | |||
| Least disadvantaged decile | 599 | 57 | 90.5 (87.8, 92.7) | 44 | 91.9 (89.2, 93.9) | 101 | 83.1 (79.9, 85.9) | |||
| Rental units | .21 | .04 | .02 | |||||||
| Most disadvantaged decile | 626 | 65 | 89.6 (87.0, 91.8) | 60 | 89.3 (86.4, 89.6) | 125 | 80.0 (76.7, 83.0) | |||
| Least disadvantaged decile | 681 | 57 | 91.6 (89.3, 93.5) | 45 | 92.8 (90.5, 94.6) | 102 | 85.0 (82.1, 87.5) | |||
| Foreign born, % | .08 | .48 | .07 | |||||||
| Most disadvantaged decile | 642 | 94 | 85.4 (82.4, 87.9) | 61 | 88.9 (85.9, 91.2) | 155 | 75.9 (72.4, 79.0) | |||
| Least disadvantaged decile | 1320 | 156 | 88.2 (86.3, 89.8) | 117 | 89.9 (88.1, 91.5) | 273 | 79.3 (77.0, 81.4) | |||
| Unemployed | .54 | .38 | .29 | |||||||
| Most disadvantaged decile | 810 | 89 | 89.0 (86.7, 91.0) | 84 | 88.4 (85.8, 90.5) | 173 | 78.6 (75.7, 81.3) | |||
| Least disadvantaged decile | 647 | 64 | 90.1 (87.5, 92.2) | 59 | 89.9 (87.1, 92.1) | 123 | 81.0 (77.7, 83.8) | |||
| Black, % | .56 | .78 | .54 | |||||||
| Most disadvantaged decile | 592 | 67 | 89.5 (88.8, 90.1) | 61 | 88.4 (85.3, 90.8) | 128 | 78.4 (74.8, 81.5) | |||
| Least disadvantaged decile | 922 | 94 | 89.8 (87.7, 91.6) | 92 | 88.9 (86.5, 90.8) | 186 | 79.8 (77.1, 82.3) | |||
Proportion of the noninstitutionalized population living below the federal poverty level.
Proportion of the population aged 18 years or older who did not graduate from high school.
Proportion of population aged 16 years or older who had operator/laborer occupations.
Proportion who spoke a language other than English at home.
Proportion of all occupied housing units with more than 1.0 persons per room.
Proportion of all occupied housing units that were renter occupied.
Proportion of population aged 16 years or older who were not employed.
| Variable | Neonatal, | Postneonatal, | Infant, AHR |
|---|---|---|---|
| Per capita income | 1.39 (0.93, 2.07) | 1.61 (1.05, 2.47) | 1.49 (1.11, 1.99) |
| Birth weight, g | |||
| <2500 | 1.37 (0.74, 2.57) | 1.13 (0.58, 2.21) | 1.28 (0.81, 2.01) |
| ≥ 2500 | 1.35 (0.79, 2.28) | 2.06 (1.18, 3.57) | 1.65 (1.13, 2.42) |
| Poverty | 1.43 (1.00, 2.06) | 1.62 (1.06, 2.47) | 1.51 (1.15, 2.00) |
| Education | 1.34 (0.93, 1.92) | 1.72 (1.18, 2.51) | 1.51 (1.16, 1.96) |
| Operator/laborer occupation | 1.53 (1.06, 2.21) | 1.54 (1.04, 2.28) | 1.54 (1.17, 2.01) |
| Non-Hispanic White | 2.09 (1.23, 3.54) | 1.45 (0.81, 2.60) | 1.78 (1.21, 2.63) |
| Non-Hispanic Black | 0.75 (0.27, 2.15) | 0.73 (0.27, 2.00) | 0.73 (0.36, 1.51) |
| Hispanic | 1.44 (0.66, 3.18) | 3.08 (1.11, 8.52) | 2.03 (1.09, 3.77) |
| Non-English speaking | 1.05 (0.78, 1.41) | 1.23 (0.89, 1.71) | 1.13 (0.91, 1.40) |
| Maternal age < 25 y | 1.10 (0.67, 1.80) | 1.16 (0.69, 1.94) | 1.12 (0.78, 1.59) |
| Maternal age 25-34 y | 0.81 (0.56, 1.32) | 1.00 (0.62, 1.62) | 0.91 (0.66, 1.26) |
| Maternal age ≥ 35 y | 1.83 (0.91, 3.68) | 3.20 (1.38, 7.39) | 2.28 (1.34, 3.87) |
| Residential crowding | 1.33 (0.86, 2.01) | 1.46 (0.96, 2.22) | 1.39 (1.04, 1.86) |
| Non-Hispanic White | 0.72 (0.38, 1.35) | 1.38 (0.72, 2.64) | 0.99 (0.63, 1.54) |
| Non-Hispanic Black | 2.43 (0.44, 13.33) | 1.16 (0.38, 3.55) | 1.42 (0.56, 3.60) |
| Hispanic | 3.99 (0.97, 16.56) | 4.42 (1.07, 18.25) | 4.24 (1.56, 11.54) |
| Foreign-born | 1.28 (0.93, 1.78) | 1.11 (0.76, 1.63) | 1.20 (0.94, 1.54) |
| Maternal age < 25 y | 1.28 (0.74, 2.21) | 0.91 (0.50, 1.63) | 1.09 (0.73, 1.62) |
| Maternal age 25-34 y | 1.20 (0.75, 1.94) | 0.84 (0.48, 1.56) | 1.05 (0.72, 1.53) |
| Maternal age ≥ 35 y | 2.22 (0.97, 5.04) | 4.24 (1.62, 11.12) | 2.75 (1.50, 5.06) |
| Unemployed | 1.37 (0.86, 1.22) | 0.90 (0.62, 1.30) | 1.10 (0.85, 1.44) |
| Rental units | 1.17 (0.76, 1.79) | 1.14 (0.72, 1.79) | 1.15 (0.84, 1.57) |
| Proportion Black | 1.05 (0.74, 1.49) | 0.95 (0.66, 1.38) | 1.00 (0.78, 1.29) |
| Proportion Hispanic | 1.07 (0.78, 1.49) | 1.44 (0.99, 2.10) | 1.22 (0.95, 1.55) |
| No. of indicators in most disadvantaged decile | |||
| 0 (Ref) | 1.00 | 1.00 | 1.00 |
| 1 | 1.24 (1.04, 1.48) | 0.94 (0.78, 1.15) | 1.10 (0.97, 1.26) |
| 2–4 | 1.33 (1.08, 1.63) | 1.00 (0.80, 1.25) | 1.15 (0.99, 1.34) |
| ≥5 | 1.36 (1.11, 1.68) | 1.17 (0.95, 1.45) | 1.25 (1.08, 1.45) |
Proportion of the noninstitutionalized population living below the federal poverty level.
Proportion of the population aged 18 years or older who did not graduate from high school.
Proportion of population aged 16 years or older who had operator/laborer occupations.
Proportion who spoke a language other than English at home.
Proportion of all occupied housing units with more than 1.0 persons per room.
Proportion of population aged 16 years or older who were not employed.
Proportion of all occupied housing units that were renter occupied.