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17-03-2017

Childhood Family Structure and Intergenerational Income Mobility in the United States

Author: Deirdre Bloome

Published in: Demography | Issue 2/2017

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Abstract

The declining prevalence of two-parent families helped increase income inequality over recent decades. Does family structure also condition how economic (dis)advantages pass from parents to children? If so, shifts in the organization of family life may contribute to enduring inequality between groups defined by childhood family structure. Using National Longitudinal Survey of Youth data, I combine parametric and nonparametric methods to reveal how family structure moderates intergenerational income mobility in the United States. I find that individuals raised outside stable two-parent homes are much more mobile than individuals from stable two-parent families. Mobility increases with the number of family transitions but does not vary with children’s time spent coresiding with both parents or stepparents conditional on a transition. However, this mobility indicates insecurity, not opportunity. Difficulties maintaining middle-class incomes create downward mobility among people raised outside stable two-parent homes. Regardless of parental income, these people are relatively likely to become low-income adults, reflecting a new form of perverse equality. People raised outside stable two-parent families are also less likely to become high-income adults than people from stable two-parent homes. Mobility differences account for about one-quarter of family-structure inequalities in income at the bottom of the income distribution and more than one-third of these inequalities at the top.

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Throughout this article, the term “both parents” denotes the two parents with whom children resided at age 0 (overwhelmingly biological parents).

A third theory holds that selection drives all mobility differences by childhood family structure. Particularly concerning are problematic traits or misfortunes generating low income among stable two-parent families (which are typically higher income) or beneficial traits or circumstances generating high income among unstable or single-parent families (which are typically lower income). In the former case, negative selection is expected to increase measured low-income persistence among children from stable two-parent families. In the latter case, positive selection is expected to increase measured high-income persistence among children from unstable or single-parent families. Following the tradition of mobility research (described in the Data and Methods section), this article aims to provide reliable population descriptions of income persistence by childhood family structure. These descriptions illuminate the rigidity of inequality; future research could isolate the causal mechanisms generating this rigidity.

My sample includes people whose childhoods span the early-1960s and early-1980s, a period of rapid family change. The share of children living with two parents dropped 11 percentage points between 1960 and 1980; between 1980 and 2016, the share dropped 7.9 percentage points (U.S. Census Bureau 2017).

When respondents were young and many lived in their parents’ households, parents provided income reports on a special survey version.

Differential measurement error across family types is unlikely to bias my results, for three reasons. First, exclusion rates due to missing income (fewer than two observations per generation) were similar between stable two-parent and other families (differing by only 3 percentage points). In general, NLSY79 has remarkably high retention and low income nonresponse rates compared with other surveys (Pergamit et al. 2001). Biases from nonrandom attrition appear inconsequential (MaCurdy et al. 1998). Second, the number of income reports contributed is also very similar across stable two-parent and other families (averaging 3.3 vs. 3.4 years in childhood and 7.1 vs. 7.5 in adulthood). Third, mobility measures are less sensitive to measurement error than might be expected (Gottschalk and Huynh 2010:311). Although classical measurement error attenuates correlations toward 0 (indicating more mobility in groups with more classical error), evidence shows that income measurement error is nonclassical. This nonclassical error often offsets attenuation biases in intertemporal correlations because errors correlate across time. These offsetting effects appear to extend beyond correlations/elasticities. Survey and administrative data produce similar earnings mobility estimates across several nonlinear measures (Dragoset and Fields 2008).

Income from nonresidential family members, including noncustodial parents, is captured through child support, alimony, and other “parental, relative support” as reported by the focal NLSY79 respondents’ parents (during childhood) or the respondents themselves (during adulthood). The survey design prevents researchers from observing other economic transfers from nonresident parents to NLSY79 respondents during childhood. The NLSY79 does capture biological parents’ education, regardless of coresidence. Consequently, it is possible to study educational mobility relative to both resident and nonresident biological parents. Prior studies have addressed this topic (e.g., Kalmijn 2015). Thus, I study income mobility.

I drop nine individuals with nonpositive incomes.

All family-structure measures capture coresidence but do not explicitly capture marital status. It is not possible to use the childhood residence calendar to separate coresidential relationships by marital status. However, this limitation is unlikely to be very problematic for this study because when the NLSY79 respondents were children (between the early-1960s and early-1980s), coresidential relationships between parents were very likely to be marital. It is also impossible to identify coresidence with “social parents” whom NLSY79 respondents did not call stepparents, adoptive parents, or foster parents in their childhood residence calendar responses. A final aspect of family complexity not captured by these data is the presence of half-siblings. This omission should not affect my conclusions because even in 2009, only 5.2 % of children lived with two biological parents and a sibling who was not a full biological sibling (Manning et al. 2014).

I include in the “stable, two parent” category 29 respondents who lived stably from ages 0 to 18 with two adoptive parents or one adoptive, one biological parent, and 11 respondents who lived stably from ages 0 to 18 with one biological parent and one stepparent. Alternate codings leave my results unchanged.

I also examined the number of years that respondents lived with both parents during different developmental stages, from ages 0–6, 7–12, and 13–18. I found no evidence that living with both parents for different periods of time matters differently for children’s income mobility if this coresidence occurs during early, middle, or late childhood.

A few children are coded as experiencing zero transitions because their residential situation is reported identically every year from ages 0–18, although their actual experience likely included transitions (e.g., children who reported living in foster care every year, or with friends). Recoding these cases as experiencing one or two transitions leaves my results unchanged.

Because incomes are logged in this canonical representation of intergenerational mobility, β measures regression to the geometric mean of adult income, not the arithmetic mean (Mitnik et al. 2015). Like the median, the geometric mean of right-skewed variables like income lies below the arithmetic mean. The geometric mean is more resistant to outliers.

I pool across genders (except when modeling earnings). I find no evidence that family income persists differently for men and women within childhood family structure groups (see also Chadwick and Solon 2002; Mitnik et al. 2015).

Without adjustment, the intergenerational income elasticity is about .15 lower among children who did not grow up in stable two-parent families than among children who did (Table 1). After adjustment using propensity score weighting, the difference is slightly attenuated, to about .11 from .15. In the weighting approach, the stable two-parent childhood family group is reweighted to capture the outcome that children from the alternative family group would have evidenced if they had (counter to fact) grown up in stable two-parent families. With an appropriate weighting model, this approach identifies causal effects under the (overly strong) assumption that conditional on the observed covariates, childhood family structure “treatment” is ignorable.

I also used Bhattacharya and Mazumder’s (2011) approach to study upward and downward rank mobility. Results confirmed the patterns evident from the multinomial and local polynomial models.

This decomposition ignores differential fertility and mortality by parental income and childhood family structure. Previous analyses have found that the contributions of these differences to recent U.S. inequality trends are relatively small (Bloome 2014; Mare 1997).

Children who lived with stepparents tended to experience more transitions than children who did not. Models including two-way and three-way interactions among parental income, childhood family composition, and number of childhood family transitions cannot distinguish differences in mobility among children who experienced two transitions but did versus did not live with stepparents.

Point estimates for two versus three or more transitions are not statistically distinguishable in the nonlinear model. Yet, neither is the difference significant between the estimate for three transitions from the linear model and the estimate for three or more transitions from the nonlinear model. The confidence interval for the nonlinear, three or more estimate is wider, reflecting data sparsity. Only about 6 % of the sample experienced three or more transitions.

Among people experiencing zero versus two childhood family transitions, 51.2 % versus 37.7 % were stably married throughout adulthood. Income mobility differs by adult family structure even though income is family size–adjusted.

Although the difference in family income elasticities between people experiencing zero versus two childhood family transitions is not statistically significant within every gender-by-adult family structure group (Table 2, panel A), F tests reveal that interactions among log parental income, number of childhood family transitions, and adult family structure can be jointly statistically distinguished from zero but cannot be distinguished from one another. These tests indicate that childhood family transitions predict family income mobility even in models that condition on adult family structure and that there is insufficient power to pinpoint how transitions predict mobility differently across adult family structures.

These hypotheses might be tested using linked census data, which should capture both spouses’ childhood family structures, assuming that married women can be linked to their parents despite surname changes.

Researchers might disagree about when to measure childhood income relative to family transitions. Future studies might explore data that permit investigations of income throughout childhood, including how mobility differs depending on the degree of homogamy among single/divorced parents who marry after a transition.

Analyses of three-way interactions among parental income, childhood family structure, and race (supported by an oversample of African American respondents in the NLSY79) indicate that both non-Hispanic white and African American children experience higher intergenerational income mobility outside stable two-parent families than within them. The mobility difference is slightly, but not statistically significantly, larger among African Americans. Yet, because African American children are much less likely than white children to grow up in stable two-parent families, the family structure–mobility association is more consequential for perpetuating income inequality among African Americans than among whites at the population level. It also contributes to the persistence of racial inequalities in income (Bloome 2014).

Aggregate income elasticities are not simple weighed averages of group-specific elasticities but also reflect income differences between groups. Consequently, decreasing the weight on stable two-parent elasticities could put downward pressure on the aggregate elasticity, but this change could be offset by changing income inequalities between family-structure groups or changing mobility patterns.

U.S. income mobility appears trendless in recent decades (Chetty et al. 2014b; Lee and Solon 2009). Forces increasing and decreasing mobility may have counterbalanced one another (Bloome 2015).

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Title: Childhood Family Structure and Intergenerational Income Mobility in the United States
Author: Deirdre Bloome
Publication date: 17-03-2017
Publisher: Springer US
Published in: Demography / Issue 2/2017
Print ISSN: 0070-3370
Electronic ISSN: 1533-7790
DOI: https://doi.org/10.1007/s13524-017-0564-4

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