Childhood Family Structure and Intergenerational Income Mobility in the United States

Abstract

The declining prevalence of two-parent families helped increment income inequality over recent decades. Does family structure also condition how economic (dis)advantages pass from parents to children? If so, shifts in the system of family unit life may contribute to enduring inequality between groups divers 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 The states. I find that individuals raised outside stable two-parent homes are much more than 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 provisional on a transition. Nonetheless, this mobility indicates insecurity, not opportunity. Difficulties maintaining middle-grade incomes create downwardly mobility among people raised outside stable two-parent homes. Regardless of parental income, these people are relatively probable to get low-income adults, reflecting a new form of perverse equality. People raised outside stable two-parent families are besides less likely to become high-income adults than people from stable two-parent homes. Mobility differences account for nearly 1-quarter of family unit-structure inequalities in income at the bottom of the income distribution and more than ane-third of these inequalities at the peak.

Notes

1. Throughout this article, the term "both parents" denotes the ii parents with whom children resided at historic period 0 (overwhelmingly biological parents).

2. A third theory holds that selection drives all mobility differences by childhood family unit construction. Particularly concerning are problematic traits or misfortunes generating low income amongst stable two-parent families (which are typically college income) or benign traits or circumstances generating high income among unstable or single-parent families (which are typically lower income). In the former case, negative option is expected to increase measured depression-income persistence amid children from stable two-parent families. In the latter case, positive pick is expected to increase measured loftier-income persistence among children from unstable or single-parent families. Following the tradition of mobility inquiry (described in the Data and Methods department), 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.

3. 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 eleven pct points between 1960 and 1980; between 1980 and 2016, the share dropped 7.9 per centum points (U.S. Demography Bureau 2017).

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

5. Differential measurement error across family unit 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 retentivity and low income nonresponse rates compared with other surveys (Pergamit et al. 2001). Biases from nonrandom compunction appear inconsequential (MaCurdy et al. 1998). 2d, the number of income reports contributed is also very similar across stable 2-parent and other families (averaging 3.3 vs. 3.4 years in childhood and 7.one vs. 7.v in adulthood). Third, mobility measures are less sensitive to measurement error than might exist expected (Gottschalk and Huynh 2010:311). Although classical measurement fault attenuates correlations toward 0 (indicating more than mobility in groups with more classical error), evidence shows that income measurement mistake is nonclassical. This nonclassical mistake 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 beyond several nonlinear measures (Dragoset and Fields 2008).

6. Income from nonresidential family members, including noncustodial parents, is captured through kid support, alimony, and other "parental, relative support" every bit reported by the focal NLSY79 respondents' parents (during childhood) or the respondents themselves (during machismo). The survey design prevents researchers from observing other economical transfers from nonresident parents to NLSY79 respondents during childhood. The NLSY79 does capture biological parents' instruction, regardless of coresidence. Consequently, it is possible to study educational mobility relative to both resident and nonresident biological parents. Prior studies take addressed this topic (e.k., Kalmijn 2015). Thus, I study income mobility.

7. I driblet nine individuals with nonpositive incomes.

8. All family unit-construction measures capture coresidence but exercise not explicitly capture marital status. It is not possible to use the childhood residence calendar to split up coresidential relationships by marital status. All the same, 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 non call stepparents, adoptive parents, or foster parents in their childhood residence calendar responses. A terminal aspect of family complexity not captured by these data is the presence of half-siblings. This omission should not affect my conclusions because fifty-fifty in 2009, only 5.two % of children lived with two biological parents and a sibling who was not a full biological sibling (Manning et al. 2014).

9. I include in the "stable, 2 parent" category 29 respondents who lived stably from ages 0 to xviii with two adoptive parents or i adoptive, i biological parent, and 11 respondents who lived stably from ages 0 to 18 with ane biological parent and i stepparent. Alternate codings leave my results unchanged.

10. I too examined the number of years that respondents lived with both parents during different developmental stages, from ages 0–6, vii–12, and 13–18. I found no prove that living with both parents for different periods of fourth dimension matters differently for children's income mobility if this coresidence occurs during early, middle, or belatedly childhood.

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

12. Because incomes are logged in this approved 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 hateful. The geometric mean is more than resistant to outliers.

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

14. Without aligning, the intergenerational income elasticity is virtually .15 lower among children who did not grow upward in stable ii-parent families than among children who did (Table ane). Later adjustment using propensity score weighting, the divergence is slightly attenuated, to near .11 from .15. In the weighting approach, the stable 2-parent childhood family grouping is reweighted to capture the outcome that children from the alternative family unit group would have evidenced if they had (counter to fact) grown upwards in stable ii-parent families. With an appropriate weighting model, this approach identifies causal effects nether the (overly strong) assumption that provisional on the observed covariates, babyhood family structure "handling" is ignorable.

15. I also used Bhattacharya and Mazumder's (2011) approach to report upward and downwardly rank mobility. Results confirmed the patterns evident from the multinomial and local polynomial models.

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

17. Children who lived with stepparents tended to feel more than transitions than children who did not. Models including 2-way and three-style interactions among parental income, babyhood family composition, and number of childhood family transitions cannot distinguish differences in mobility amid children who experienced two transitions but did versus did non live with stepparents.

18. Signal estimates for ii versus three or more transitions are not statistically distinguishable in the nonlinear model. Yet, neither is the difference meaning between the guess for three transitions from the linear model and the estimate for iii or more transitions from the nonlinear model. The confidence interval for the nonlinear, three or more gauge is wider, reflecting data sparsity. Simply about six % of the sample experienced three or more transitions.

19. Among people experiencing zero versus two babyhood family transitions, 51.ii % versus 37.7 % were stably married throughout adulthood. Income mobility differs by adult family unit structure even though income is family size–adjusted.

20. Although the deviation in family income elasticities betwixt people experiencing zero versus 2 childhood family transitions is not statistically significant inside every gender-by-adult family construction group (Table 2, panel A), F tests reveal that interactions among log parental income, number of childhood family transitions, and adult family unit structure can be jointly statistically distinguished from zip but cannot exist distinguished from one some other. These tests indicate that childhood family unit transitions predict family income mobility fifty-fifty in models that condition on developed family structure and that there is bereft power to pinpoint how transitions predict mobility differently across adult family structures.

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

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

23. Analyses of iii-manner interactions among parental income, childhood family structure, and race (supported past an oversample of African American respondents in the NLSY79) indicate that both not-Hispanic white and African American children experience higher intergenerational income mobility outside stable two-parent families than inside them. The mobility divergence is slightly, only not statistically significantly, larger amidst African Americans. Withal, because African American children are much less probable than white children to grow up in stable two-parent families, the family structure–mobility association is more consequential for perpetuating income inequality amongst African Americans than among whites at the population level. Information technology besides contributes to the persistence of racial inequalities in income (Bloome 2014).

24. Aggregate income elasticities are non simple weighed averages of group-specific elasticities only also reflect income differences betwixt groups. Consequently, decreasing the weight on stable ii-parent elasticities could put downwardly force per unit area on the aggregate elasticity, just this change could be offset by changing income inequalities between family-structure groups or changing mobility patterns.

25. U.South. 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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Acknowledgments

This article has benefitted from the useful comments of Census editors and reviewers, Paula Fomby, Christopher Jencks, Karen Lacy, Laura Tach, Bruce Western, William Julius Wilson, Lawrence Wu, and seminar participants at Columbia University and the Academy of Pennsylvania. This inquiry was supported by a Ford Fund grant from the Population Studies Center at the University of Michigan and an NICHD center grant to the Population Studies Heart at the University of Michigan (R24 HD041028).

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1. Department of Sociology, Population Studies Center and Survey Research Heart, University of Michigan, 426 Thompson Street, Ann Arbor, MI, 48104, U.s.

   Deirdre Bloome

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Correspondence to Deirdre Bloome.

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Bloome, D. Childhood Family Structure and Intergenerational Income Mobility in the United States. Demography 54, 541–569 (2017). https://doi.org/10.1007/s13524-017-0564-4

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• Published: 17 March 2017

• Issue Engagement: April 2017

• DOI : https://doi.org/10.1007/s13524-017-0564-4

Keywords

• Intergenerational mobility
• Family structure
• Family dynamics
• Income inequality

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