racial differences in the hurdling of prenatal care barriers

R A C I A L D I F F E R E N C E S IN T H E H U R D L I N G O F

PRENATAL CARE BARRIERS

Geoffrey Warner

The success of black or white mothers in obtaining adequate prenatal care is examined. Two departures from public health convention are employed. The independent variables' marginal effects are calculated from their logit coefficients. The odds ratio of care adequacy between races is derived from race-specific regressions. It yields a smaller variance and type II decision error likelihood compared to the race dummy method.

A working-class life outlook and apathetic fathers are the highest barriers to adequate care. Wantedness, in the form of desired timing, is a very strong motivator. Improving upward socioeconomic mobility and paternal attitudes are important aspects of increasing prenatal care adequacy rates.

I N T R O D U C T I O N

Prenatal care receives a tremendous amount of attention in the public health literature. The focus on the absence of or the late initiation of care is well deserved. There is a much higher proportion of low weight births among mothers receiving no or late care than among mothers receiving adequate care (Kotelchuk, 1994b). Public health professionals have investigated and written extensively about the barriers to prenatal care in an effort to identify interventions aimed at improving the utilization of care, especially among poor and minority women. Economists have not examined this issue with appreciable systematic rigor, even though the problem lends itself to the market entry analysis of the reservation wage in labor market entry or the reservation price in home purchases, partly due to insufficiently rich data. This article intends to start to fill a large void by examining, from an economic perspective, the factors contribut- ing to a mother's success in obtaining adequate prenatal care

The article will add to the prenatal care literature in three ways. The

96 The Review of Black Political Economy/Winter 1997

adequacy of care will be measured on two dimensions; timing adequacy and visits adequacy. This is to accommodate results indicating that factors affecting timing of care initiation are not identical to the factors affecting the frequency of visits (Oberg et al., 1991). Logistic regressions are performed on a dichotomous adequate versus inadequate care variable. But, breaking from public health tradition, they yield marginal effects, the percentage point change in the likelihood of obtaining adequate care due to a one unit change in an independent variable, rather than the usual odds ratios. Unlike an odds ratio, which conveys the magnitude of a variable's influence relative to that of a base variable, the marginal effects method conveys the absolute magnitude of a variable's influence on the likelihood of obtaining adequate care. Like Cooney (1985) and Joyce (1994), analyses are performed on blacks and whites separately, instead of the traditional combined regression with race dummies, (for example Rosenweig and Schultz [1991]). There are two reasons for this. One, it permits the slope coefficients to vary across races. This allows the comparison and testing of racial differences in the contributions of a woman's characteristics to the attainment of adequate prenatal care. Two, by calculating the difference between index function values, an odds ratio between mothers of different races can be calculated with a smaller variance than can be achieved using race dummies in a combined regression.

The following sections describe the model, the data set and the estimation methods, and the results. The final section outlines some policy implications.

THEORETICAL MODEL

A modest review of the literature yielded some articles (Lia-Hoagberg et al., 1990; Laviest, Keith and Gutierrez, 1995; Rogers and Schiff, 1996) that share the framework under which economic analysis takes place. The authors investigated not only factors which inhibit access to care but also factors which enhance access to care. This is the same approach of Herzberg's Two Factor theory in organizational behavior on the issue of job satisfaction. The hypothesis is that there are two independent groups of factors. One, the hygienes, affect how much one dislikes one's job; the other, the motivators, affect how much one likes one's job (Hellriegel and Slocum, 1979, pp. 408-10). Reducing how much a worker dislikes his job may not affect how much he likes it. The analogous economic approach is that there are costs to care which discourage its use--the

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barriers; and there are benefits to care which encourage its use-- the motivators or enablers. When the benefits to care exceed its costs, care is obtained; when its costs exceed its benefits, care is not obtained. The key point is that lowering the costs or barriers of prenatal care does not, by direct consequence, increase the motivation to obtain prenatal care and therefore does not necessarily increase care utilization.

The above concept sets up the theoretical framework for this analysis. Every mother is assumed to have the ability to determine, when she makes some effort to obtain prenatal care, how much additional benefits and additional costs her effort will yield or generate. The researcher cannot observe this net benefit to effort. What is observed is whether the mother obtains care or not. However, the researcher knows that if care is (is not) obtained, the mother's net benefits must have been positive (negative). The underlying model is

I* = ~'X + e

where I* is the unobserved net benefit of effort to obtain prenatal care and X are the independent variables. The model to be estimated by logit is

Y = I 3 ' X + e

where Y = 1 if I* > 0

Y = 0 i f I * <- -0 .

This regression will yield the probability that a mother's efforts were sufficient to obtain adequate care, given her characteristics. The sign of a coefficient will reveal whether that characteristic helps or hinders a mother in her efforts to obtain care.

There are four groups of explanatory variables. The variables used to measure income are total household income, source of income, mother's and father's education, and method of financing prenatal care. The variables that measure availability of care are rural residence and state level per capita health care facilities and professionals. The variables that capture need are mother's age, parity, and number of children in the household. The variable that represents motivation is degree of wantedness. (Other included variables are described in Appendix, Table A1.)

The dependent variable measures care adequacy. Care initiation is considered adequate if it occurs in the first trimester. 1 Care utilization is considered adequate if 80 percent or more of the ACOG recommended


visits are received. 2 Since actual visits made depends on gestation, the comparison is to the number of visits recommended for the gestation of that birth)

This framework does have its weaknesses. The most serious is that a particular net benefit is not determined by a unique cost/benefit pair. A highly motivated mother, a mother who perceives high benefits to her effort to obtain care, facing high barriers may have the same net benefit to effort as a moderately motivated mother facing moderate barriers. This lack of one-to-one correspondence between cost/benefit pairs and net benefits introduces an element of heterogeneity into the characteristics of mothers who obtain or do not obtain adequate prenatal care. Not obtaining adequate care is not necessarily indicative of high barriers to care, it may also indicate low motivation to obtain care. Most of the research on prenatal care access focuses on barriers (Aved et al., 1993; Boggs et al., 1995; Curry, 1989; Perez-Woods, 1990; Poland et al., 1987; Joyce et al., 1983; Johnson, Primas and Coe, 1994; Institute of Medicine, 1988; Meikle et al., 1995; Sable et al., 1990). Implicit in the results of such studies is the assumption that lowering barriers to care will increase its utilization. However, lowering access barriers to care for women who are not motivated to seek care will be ineffective.

I have tried to address this weakness by including a measure of motivation; the wantedness of the pregnancy. This permits the analysis to vary the barriers to care while holding the motivation, wantedness, constant. Now, moderate motivation/high barriers can be distinguished from low motivation/high barriers, and high motivation/low barriers from moderate motivation/low barriers. In turn, this helps distinguish between a high motivation/high barrier net benefit and a low motivation/low barrier net benefit. It is unrealistic to expect the wantedness variable by itself to capture all the aspects of motivation to obtain prenatal care. Conse- quently, this article, while not making the same assumptions about motivation as other articles, still suffers from this heterogeneity weakness but to a lesser degree.

A second weakness relates to the methodology. The mothers are sepa- rated into racial groups for analysis purposes. The very act of grouping observations into race-specific samples can affect the barriers and benefits implicit in the regressions. Blacks and whites may each, as a group, face different barriers to prenatal care for political, sociological, and cultural reasons that are not captured by the included variables. For example, some medical staff may, consciously or unconsciously, make moral or value judgements about single pregnant women that discourage future

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contact with the health care system. As a larger proportion of black mothers than white mothers are single, such behavior would disproportion- ately affect black women. This violates the ceteris paribus assumption that is needed for meaningful comparisons across regressions, even though the coefficients are statistically consistent in each individual regression.

A greater weakness lies in the combined regression because it forces equality of black and white slope coefficients without addressing race- specific barriers. Reducing the effect of racial differences in the process of obtaining adequate prenatal care to a shift in the intercept, masks important details of that process on a race-specific level. The race-specific regressions yield much richer results, highly descriptive of the process unique to each race. This is a strength that far outweighs the weakness described in the previous paragraph if one is more interested in understanding each race individually than in understanding their similarities and differences.

Despite these weaknesses, the results of this research illuminate a not well considered path toward an improvement in infant health outcomes. That illumination is generated, in part, by a different method of calculating odds ratios. The odds ratio calculated from the race dummy coefficient in a combined regression can be misleading because it is constant over the entire range of values of the explanatory variables. Using a methodology borrowed from labor economics, which permits the odds ratio to vary with the characteristics of the women involved, enables the investigator to obtain a better understanding of why prenatal care adequacy differs across races. The race specific regressions used in this analysis permit the index value difference (the difference in log odds) to vary across races as the mother's characteristics vary, rather than be constant. Furthermore, the variability of the index value difference is from two sources. One source is the values of the explanatory variables, i.e., the maternal characteristics. The other source is the coefficients, i.e., the contribution each variable makes toward obtaining adequate care, which represents the conversion of maternal characteristics into prenatal care adequacy. Thus, the predicted difference in the likelihood of attain- ing adequate prenatal care can be broken down into two components; differences in observed characteristics and differences in the likelihood of contributions of those characteristics.

The characteristics component can be isolated by calculating two index values using the coefficients of one race and the mean values of the explanatory variables of both races. When the exponential constant e is raised to the difference of the two index values, the result is the odds


ratio of care adequacy due solely to differing maternal characteristics. The coefficients component can be isolated by again calculating two index values using the means of one race and the coefficients of both races. This will yield an odds ratio of care adequacy due solely to differing coefficients. The two components have very different policy implications. A large odds ratio due to maternal characteristics implies a policy directed at mothers. A large odds ratio due to the coefficients implies a policy directed at the system or process in which the mothers are func- tioning, where the regression coefficients are considered descriptive of the system or process which converts maternal characteristics into care adequacy and not descriptive of the mother.

DATA AND ESTIMATION

The data come from the National Maternal and Infant Health Survey (NMIHS) of 1988. This is a sample of all conception outcomes (live birth, fetal death, infant death) occurring in 1988 in 48 states and the District of Columbia. (Montana and South Dakota declined to participate in the survey.) The survey was intended to explore the causes of poor birth outcomes so low weight births and minority births were over- sampled. Observation weights for obtaining population estimates are available and used in the regressions, meaning that weighted logit regressions were performed. The NMIHS data was supplemented by data on the state of maternal residence from the State and Metropolitan Area Data Book 1991. This area data was mostly for 1988, the population data was from 1990. The age of the data is a weakness, especially given several states' Medicaid reform efforts since 1988. Detailed policy-specific conclusions cannot be drawn from the results, but conclusions regarding general policy directions are still valid.

Single live births to non-Hispanic black and non-Hispanic white women aged at least twenty-one years comprised the universe. To reduce the inclusion of incorrectly recorded data, observations with gestation less than twenty or more than forty-five weeks, or birthweights below 400 grams or above 6,000 grams were excluded.

Teenagers are excluded from the analysis because any behavioral differences between teenage and adult mothers and their impact on care adequacy are better examined separately from the general examination of prenatal care adequacy. An adolescent's level of education and marital status may be significantly correlated with the state of pregnancy. This issue of endogeneity of education and marital status is curtailed by ex- cluding adolescents from the analysis.

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Many variables had missing values for some observations, which were imputed by the National Center for Health Statistics (NCHS). The imputation method was to find another complete observation with the most similar maternal characteristics and use that value for the missing one. Few imputation rates exceeded five percent. 4

The logit coefficients, their t-scores, and marginal effects are taken directly from the statistical output. The marginal effects are calculated by multiplying the coefficients by the attenuation factor value. The attenuation factor value is the logistic density evaluated at a particular point. That point in this article is the (weighted) mean values of the explanatory variables, displayed in the Appendix, Table A2. 5

To compute the odds ratios, the appropriate index values (coefficients times variable means) are first calculated. Their variances are calculated by pre- and post-multiplying the coefficient covariance matrix by the vector of mean values. The results of these calculations are displayed in Table 5. Treating each cell of the table as a random variable with its variance, the standard difference between two random variables' calcula- tion will yield odds ratios and their associated confidence intervals.

RESULTS

Almost all of the coefficients in the care initiation adequacy equation (Tables 1-2), which are significant at 5 percent, have the expected sign. The exceptions are Medicaid financing, number of children in the household, and foreign birth.

The payment method coefficients contain an anomaly. In the black regressions the self-pay coefficient is smaller than the Medicaid coefficient. This is to be expected because Medicaid recipients face a lower money price of prenatal care than do self-payers and therefore would use more of it, all else being equal. In the white regression, however, the self-pay coefficient is larger than the Medicaid coefficient (zero versus a significant negative value). McDonald and Cobum (1988) found the same result in a sample of almost 4,000 predominantly white Maine mothers. This implies that, ceteris paribus, white Medicaid recipients are less likely to start prenatal care in the first trimester than are white self-payers even though they face a lower money price of care than do self-payers.

As the number of children in a household increases, one can expect greater demands on the mother's time, assuming that she is the primary caretaker of the household's children, and in turn, longer delay. This expectation holds for whites. The coefficient on the number of children is


TABLE 1 Care Initiation Adequacy (Whites)

variable (e-2) coefficient Itl marginal effect

Intercept 30.14 0.39 1.41

Wanted Earlier -23.38 1.22 -1.09 Wanted Later -87.19 5.50 -4.07 Never Wanted -112.11 4.81 -5.24 Parity -13.63 1.38 -0.64

Mother's Education 2.59 0.70 0.12 Age 2.28 1.19 0.11

Father's Education 6.90 2.20 0.32 Age -1.53 1.15 -0.07

Black Father -71.78 1.88 -3.35 Hispanic Father -52,02 2.15 -2.43 Father Other Ethn. -11.67 0.28 -0.55

Household Income 1,19 2.64 0.06 Mother Works -1,05 0.08 -0.05 Number of Child. -17.88 1.89 -0.84

Self-pay 1.09 0.08 0.05 Medicaid --68.29 3.42 -3.19 Other -27.10 1.02 -1.27

Aid to Poor 22.05 1.16 1.03 SSI / Vet -26,03 0.93 -1,22 Other Sup, -4.51 0.21 -0.21

Stable Res, 47.46 3.34 2.22 Rural Res. -29.47 2.16 -1.38 Single mother -34.19 1.73 -1.60 Foreign Born -27.44 0.82 -1.28

Physicians per capita 0.38 2.32 0.02 Health Exp. per capita 2.00 1.18 0.09

Cohabitants-father 47.89 1.35 2.24 mother' s parents -24.41 0.61 - 1.14 alone with child, 53.40 1.26 2.50 other 81.20 1.59 3.79

Chi-sq value (df=30) 386.92 (p = 0.00001).

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TABLE 2 Care Initiation Adequacy (Blacks)

variable (e-2) coefficient Itl marginal effect

Intercept -52.93 0.99 -7.56

Wanted Earlier -26.08 1.55 -3.73 Wanted Later -74.31 6.45 -10.62 Never Wanted -89.32 6.47 - 12.76 Parity -20.23 4.63 -2.89

Mother' s Education 9.59 3.73 1.37 Age 2.98 2.51 0.435

Father's Education -0.18 0.08 -0.03 Age 0.37 0.48 0.05

White Father - 132.21 4.56 - 18.89 Hispanic Father -13.36 0.48 -1.91 Father Other Ethn. -71.27 0.76 -10.18

Household Income 1.62 4.49 0.23 Mother Works 15.38 1.60 2.20 Number of Child. 7.81 2.16 1.12

Self-pay 57.88 4.49 8.27 Medicaid 75.91 6.73 10.85 Other 17.97 0.86 2.57

Aid to Poor -1.00 0.09 -0.14 SSI / Vet -7.92 0.56 -1.13 Other Sup. -15.03 1.20 -2.15

Stable Res. 12.08 1.17 1.73 Rural Res. -16.71 1.59 -2.39 Single mother -17.09 1.59 -2.44 Foreign Born 75.67 3.13 10.81

Physicians per capita -0.01 0.09 -0.01 Health Exp. per capita 0.64 0.65 0.09

Cohabitants-father 14.78 0.73 2.11 mother's parents --47.37 2.33 -6.77 alone with child. 1.75 0.08 0.25 other cohabs -26.98 1.51 -3.85

Chi-sq value (df=30) 383.20 (p = 0.00001),


significantly negative. The opposite is observed for blacks. The number of children coefficient is significantly positive in that regression. I have no non-speculative explanation for this result.

The foreign maternal birth coefficient is unexpectedly positive in the black regression. Foreign birth makes a black woman 2.13 times more likely (95 percent CI [1.33, 3.42]) than an otherwise identical native born black woman to start prenatal care in the first trimester. Foreign birth has no effect on the white likelihood of adequate care initiation. I have no non-speculative explanation for this result.

Notwithstanding the racial difference in unobserved barriers, it is help- ful to point out some coefficient differences. The coefficients on mother 's age and mother 's education are both significantly positive in the black regression but insignificant in the white regression. Parity is significantly negative in the black regression but insignificant in the white regression. The father's education coefficient is significantly positive in the white regression but insignificant in the black regression. The physicians per capita coefficient is significantly positive in the white regression but insignificant in the black regression. A black mother's personal characteristics seem to have a greater impact on her care initiation adequacy than those of a white mother on her care initiation adequacy. However, aspects of a white mother's environment (e.g., father's and state's characteristics in the form of education and physician density) seem to have a greater impact on her care initiation adequacy than those of a black mother 's environment on her care initiation adequacy. My interpretation is that society helps see to it that white women start prenatal care early, whereas black women must accomplish it on their own.

Almost all of the coefficients in the care utilization adequacy equation (Tables 3-4), which are significant at 5 percent, have the expected sign. The exceptions are cohabiting with maternal parents and employed mother.

Mothers who live with their parents without the father present are less likely to receive an adequate number of visits than a childless pregnant woman living alone. What might be occurring is that the women living alone are making extra visits to obtain support that these other women are getting from their parents. Mothers living with their parents may be substituting some parental care for physician or nursing care. All of the cohabiting coefficients are negative, further suggesting that persons living with a pregnant woman provide or do things that a solo pregnant woman is obtaining from the health care system.

Working decreases the amount of leisure time and increases the price of time. This should result in prenatal care visits becoming more costly

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TABLE 3 Care Utilization Adequacy (Whites)

variable (e - 2) coefficient Itl marginal effect

Intercept 46.07 1.01 10,20

Wanted Earlier 2.03 0.24 0.45 Wanted Later -10.68 1.27 -2.36 Never Wanted - 19.03 1.22 --4,21 Parity -8.85 1.47 -1.96

Mother's Education 2.86 1.45 0.63 Age 0.21 0.19 0.05

Father's Education -2.75 1.59 --0.61 Age 0.12 0.15 0.03

Black Father 16.51 0.52 3.66 Hispanic Father 4.20 0.25 0.93 Father Other Ethn. 64.91 2.20 14.37

Household Income 0.34 1.62 0.08 Mother Works 3.69 0.47 0.82 Number of Child. -10.83 1.85 -2.40

Self-pay 11.40 1.62 2.52 Medicaid -5.60 0.41 -1.24 Other 28.98 1.72 6.42

Aid to Poor 10,45 0.81 2.31 SSI / Vet 12.21 0.58 2.70 Other Sup. 9.79 0.76 2.17

Stable Res. 12,78 1,45 2.83 Rural Res. 7.11 0.88 1.57 Single mother -34.33 2.50 -7.60 Foreign Born -12.68 0.72 -2.81


Cohabitants-father -45,15 1.62 - I 0.00 mother's parents -63.30 1.87 -14.01 alone with child. -30.90 0.91 -6.84 other cohabs -9.37 0.23 -2.07

Chi-sq value (df=30) 112.35 (p = 0.00001),


TABLE 4 Care Utilization Adequacy (Blacks)

variable (e - 2) coefficient Itl marginal effect

Intercept -188.97 4.58 -46.96

Wanted Earlier 11.50 1.01 2.86 Wanted Later -8.90 1.10 -2.21 Never Wanted -26.80 2.51 --6.66 Parity -15.71 4.27 -3.90

Mother' s Education 4.38 2.20 1.09 Age 2.59 2.79 0.64

Father's Education 4.05 2.27 1.00 Age 1.03 1.67 0.26

White Father -36.40 1.32 -9.04 Hispanic Father 0.59 0.03 0.15 Father Other Ethn. 98.75 1.12 24.54

Household Income 0.77 3.09 0.19 Mother Works 18.76 2.43 4.67 Number of Child. 3.09 1.02 0.77

Self-pay 15.97 1.74 3.97 Medicaid 29.55 3.25 7.34 Other 4.91 0.30 1.22

Aid to Poor 12.19 1.40 3.03 SSI / Vet 17.53 1.46 4.36 Other Sup. 6.81 0.64 1.69

Stable Res. -0.83 0.10 -0.21 Rural Res. -6.28 0.72 -1.56 Single mother -38.45 4.63 -9.56 Foreign Born -37.72 2.52 -9.37


Cohabitants-father -17.48 1.08 -3.35 mother' s parents --40.42 2.41 - 10.04 alone with child. -17.70 1.03 --4.40 other -43.40 2.12 -10.79

Chi-sq value (df=30) 252.51 (p = 0.00001).

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and fewer of them made. However, the work coefficient in both regressions is positive, significantly so in the black regression. There are two possible explanations for this. Unobserved characteristics related to em- ployability may be correlated to the perception of the need for prenatal care. Employment provides income and, for full-time employees, health benefits. For black mothers, the income effect from working may domi- nate the substitution effect. The growth of managed care also needs to be considered. If working mothers are more likely to be in a managed care environment than nonworking mothers then the greater emphasis on pre- vention, relative to fee-for-service care, may yield the observation of greater likelihood of visits adequacy for employed mothers.

The foreign maternal birth coefficient is significantly negative in the black regression, as expected. It is curious how this variable affects care initiation and care utilization in opposite ways for blacks while it has no effect on either for whites.

As in the care initiation regressions, the coefficients in the care utilization regressions on mother's age and education are significantly positive and the parity coefficient is significantly negative for blacks but all are insignificant for whites. The Medicaid coefficient is significantly positive for blacks but insignificant for whites, similar to the care initiation regression. In contrast to the initiation regressions, the father's education and physician density coefficients for care utilization are significantly positive in the black regression. The environment will help a black woman stay in prenatal care, once she has started.

The coefficients on pregnancy wantedness are intriguing. Strong similarities are evident in the two samples. The coefficients on the variables for wanting the infant later or never in the care initiation regression are large and highly significant. In the white regression they are the largest of all the significant coefficients, indicating that inopportune timing has the largest impact on the adequacy of prenatal care initiation of all the observed variables. In the black regression they also have the largest significant coefficients after that on white father. The small proportion of white fathers in the blacks sample (1.4 percent) permits me to state that, for blacks also, inopportune timing has the largest impact on the adequacy of prenatal initiation of all the observed variables. The most important difference is that black women reported "too soon" pregnancy almost twice as much as did white women. In addition, the marginal effects of pregnancy wantedness are twice as large for blacks as for whites. Opportune timing has much greater intensive and extensive impacts on care initiation adequacy for blacks than for whites. The coef-


ficients on wantedness in the care utilization regressions are all insignificant in the white sample. Only the coefficient on the never wanted variable is significant in the black sample and its magnitude is less than that of several other coefficients. Wantedness has a very limited impact on the likelihood of achieving adequate care utilization.

Lia-Hoagberg et al.'s (1990, p. 492) research on motivators for prenatal care indicates that the most frequently given reason for seeking prenatal care was the desire for a healthy baby (adequate care 52 percent, intermediate care 48 percent, inadequate care 27 percent). The most frequently given reason for seeking prenatal care by women who received inadequate care was that they felt they were supposed or expected to do it, (28 percent). The results of this article are consistent with their findings. The wantedness coefficients in the initiation adequacy regressions indicate that the desire for a baby now, as opposed to later or not at all, is a strong indicator of the regard in which the mother holds her baby's health. Mothers appear to have a low regard for the health of unwanted babies which manifests itself more as long delay in starting care than as a small number of care visits. Of course, pregnancy termination is an ex- treme manifestation of unwantedness. (Joyce and Grossman [1990] investigated the choice of carrying a pregnancy to term as a manifestation of wantedness.)

Lia-Hoagberg et al. (1990, p. 492) also identified what they termed supports for prenatal care. Most often this support was in the form of a person who encouraged or helped the mother obtain care. White women (43 percent) were significantly more likely to have received such support from a male (husband or boyfriend) than were black women (23 percent) who most frequently received support from their mothers. Their findings are consistent with the significant and large coefficient found on the single mother variable in the utilization adequacy regression.

The results of the odds ratio calculations are not surprising. Whites are four times more likely than blacks to start care in the first trimester (OR = 4.03, 95 percent CI [3.36, 4.83]) and 74 percent more likely to receive at least 80 percent of the ACOG recommended number of visits (OR = 1.74, 95 percent CI [1.59, 1.90]). Along a row in Table 5, coefficients are compared while holding means constant; thus the contribution of the coefficients is isolated. In a column of the same table, means are compared while holding the coefficients constant, thus the contribution of maternal characteristics is isolated.

Comparing coefficients using black mean values, the odds ratio for first trimester care is 1.26 (95 percent CI [1.03, 1.55]). A white with

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black mean values is 26 percent more likely to start care in the first trimester than a black with black mean values. Comparing coefficients using white mean values, the odds ratio for first trimester care is 2.03 (95 percent CI [1.61, 2.58]). A white with white mean values is twice as likely as a black with white mean values to obtain first trimester care. The interesting finding here is the dramatic difference in the effect of white versus black coefficients. This result expressed from a black perspective says that a woman with white mean values (say middle class) is twice as likely to start care in the first trimester if she uses the white system or process of care compared to using the black system or process of care.

Comparing mean values using black coefficients, the odds ratio for first trimester care is 1.98 (95 percent CI [1.63, 2.41]). A black with white mean values is twice as likely to start care in the first trimester as a black with black mean values. Comparing mean values using white coefficients, the odds ratio for first trimester care is 3.19 (95 percent CI [2.50, 4.07]). A white with white mean values is more than three times as likely as a white with black mean values to obtain first trimester care. Here we see the strong influence of maternal characteristics on care initiation. It suggests that improving black maternal characteristics (middle class vs. working class) can be a highly effective means of improving first trimester care rates. I speculate that delaying pregnancy to a more desirable time, affords black women the opportunity to complete educations and earn higher incomes, accumulate savings, and find supportive mates or husbands, in other words, to take on middle class characteristics.

Comparing coefficients using black mean values, the odds ratio for 80 percent care utilization is 1.22 (95 percent CI [1.05, 1.42]). A white with black mean values is 20 percent more likely to obtain an adequate number of visits than a black with black mean values. Comparing coefficients using white mean values, the odds ratio for care utilization adequacy is 1.06 (95 percent CI [0.93, 1.21]). A white with white mean values is no more likely than a black with white mean values to achieve care utilization adequacy. The noteworthy finding here is that, for middle class women, the effect of the difference in systems or processes of care is much less for care utilization than for care initiation. Black middle class women have less difficulty converting their middle class characteristics into adequate care utilization than into adequate care initiation.

Comparing mean values using black coefficients, the odds ratio for 80 percent care utilization is 1.63 (95 percent CI [1.44, 1.86]). A black with white mean values is 60 percent more likely to obtain adequate care utilization than a black with black mean values. Comparing mean values


using white coefficients, the odds ratio is 1.42 (95 percent CI [1.22, 1.65]). Whites with white mean values are 42 percent more likely than whites with black mean values to obtain 80 percent utilization. What is noteworthy is the modest effect the difference in maternal characteristics has on the likelihood of care utilization adequacy compared to care initiation adequacy.

The main importance of these results lies in the decomposition of the total odds ratio. Comparisons of the index values along the left to right diagonal of Table 5 yield the total odds ratio. Comparisons of the index values along the two sides of the triangle formed with the diagonal yield the decomposition into characteristics effect and coefficients effect. The decomposition for care initiation indicates that the total odds ratio is dominated by the characteristics effect. The interpretation is that improving socioeconomic characteristics is a more effective way of improving the rate of first trimester care than is intervening in the system, particularly for the disadvantaged. Intervening in the system should be inter- preted as trying to affect the regression coefficients, trying to change them from the black values to the white values. The decomposition for care utilization yields similar results. The characteristics effect dominates the coefficients effect. Again the interpretation is that improving socioeconomic characteristics will improve the rate of care utilization more than will intervening in the system.

A finding that cannot be ignored is that the care initiation odds ratio shows little racial difference between disadvantaged blacks and whites but does show a significant racial difference between advantaged blacks and whites. It is disturbing that their respective systems yield equally poor prenatal care initiation adequacy rates for working class blacks and whites, and that middle class blacks should have poorer prenatal care initiation adequacy rates than middle class whites. It suggests that there are class-related barriers to adequate prenatal care among working class mothers, and race-related barriers to adequate prenatal care among middle class mothers. The overrepresentation of blacks in the working class indicates that it is imperative to identify and distinguish between class- related issues and race-related issues.

POLICY IMPLICATIONS

Two main policy implications are evident. Inopportune timing of the pregnancy is a very high barrier to care adequacy, and particularly so for care initiation. The motivation of all mothers to obtain adequate prenatal

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TABLE5 ~ d e x V a l u ~ a n d V a ~ a n c e s

coefficients Black White means

Care Utilization

Black

White

Carelnitiation

Black

White

0.1529 0.3537 t.032e-3 4.963e-3

0.6431 0.7044 3.294e-3 1.072e-3

1.569 1.804 2.063e-3 9.025e-3

2.253 2.963 8.034e-3 6.487e-3

Note: ae-b = a xl0 -b. Variances are below the index values. The above figures cannot be reproduced by the reader because the coefficient covariance matrix has not been reported. It is available from the author upon request.

care would be increased by improving the timing of their pregnancies. Black women would benefit greatly from this because they seem to have more difficulty than white women in achieving the desired pregnancy timing. Such an intervention would include family planning efforts that enable women to time their pregnancies to better match their tastes and economic resources.

The second policy implication relates to socioeconomics. The odds ratio decomposition strongly suggests that improving the socioeconomic well-being of working-class mothers is an effective method of increasing prenatal care adequacy, care initiation in particular. The coefficient on the household income variable strongly suggests that increasing income will increase care adequacy, particularly care initiation adequacy. The consistently insignificant coefficients on the source of income variables are a strong suggestion that the composition of household income is much less relevant than its amount. Taken together, the results say that improving socioeconomic well-being through income maintenance programs can increase prenatal care adequacy rates. However, the maternal education coefficient implies that educational success, and the greater confidence and desire for self-determination that comes with greater


knowledge and reasoning ability, is also an important factor for black women.

The consistently negative coefficients on the single mother variable and the positive coefficients on the cohabiting father variable in the care initiation regressions suggest that uninvolved fathers are a high barrier to adequate prenatal care, particularly care initiation adequacy. Paternal attitudes and behaviors towards a pregnancy may be an especially important factor if maternal wantedness is related to them.

I recognize that improving socioeconomic well-being is a long-term task. Interventions to make the care system more accessible to the cur- rently disadvantaged are still necessary, so Medicaid expansion plans should be supported (Braverman et al., 1993; Coburn and McDonald, 1992). Interventions designed to improve mate selection and change attitudes toward paternal responsibilities will also take years to generate significant results. However, the only way to achieve a substantial and sustained increase in prenatal care adequacy is to improve the economic well-being of the population in general, mothers in particular, and improve attitudes towards pair bonding and paternal obligations. The minority community has much to gain from such interventions. Therefore, its leaders should consider being the vanguard in conceptualizing and implementing programs designed to explain to adolescents, boys as well as girls, why the professional monitoring of a pregnancy and the sustained involvement of the father are so important to the delivery of a healthy child.

NOTES

1. The adequacy standard for care initiation is taken from the Kessner Adequacy of Prenatal Care Index (Kessner et al., 1973). It states that a necessary condition for care adequacy is a first trimester initiation and a delivery by a private obstetrician. The private delivery standard was dropped for this article.

2. ACOG recommends that a pregnant woman visit her obstetrician once every 4 weeks from conception until week 28, then once every two to three weeks until week 36, and once every week thereafter until birth (American College of Obstetrics and Gynecology, 1988). This leads to a recommendation of 15 visits for a full-term pregnancy (forty weeks), 80 percent of which is 12.

3. Milton Kotelchuk (1994a) has developed a visits measure that adjusts for the timing of the first prenatal care visit as well as the timing of the birth. I reject its use in this article because it captures the intensity of prenatal care use much more than it does the quantity of use. His threshold of adequacy of received visits of 80 percent of the ACOG recommendation is used here.

4. The high imputation rate variables for whites are maternal work week (7.6 percent) and household income (10.5 percent). The high imputation rate variables for

Warner 113

blacks are mother works (12.9 percent), father's education (8.7 percent), Hispanic father (7.1 percent), and household income (18.0 percent).

5. The marginal effects will have the same t-scores as the coefficients because the attenuation factor, evaluated at a point, is a constant. These calculations are explained by Green (1993, pp. 639-48).

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racial differences in the hurdling of prenatal care barriers

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