Ed 1000 Attainment in Social Edu

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Effects of Family Structure on the Earnings Attainment Process: Differences by Gender

Author(s): Mary Ann Powell and Toby L. ParcelSource: Journal of Marriage and Family, Vol. 59, No. 2 (May, 1997), pp. 419-433Published by: National Council on Family RelationsNational Council on Family RelationsStable URL: http://www.jstor.org/stable/353480

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MARY ANN POWELL AND TOBY L. PARCELThe Ohio State University

Effects of Family Structureon the EarningsAttainmentProcess: Differences by Gender

This study compares how being raised in an orig-inal, two-parent family and being raised in otherfamily structures affects educational achievement,occupational status, and earnings attainmentfora national sample of 30- to 59-year-old womenand men. Data are derived from the 1989 PanelStudy of Income Dynamics. Findings suggest thatfamily structure has different effects by gender.Although both men and women from original,two-parent families earn more, on average, thanthose from other family structures, for women,this effect occurs through educational attainment.For men, the association betweenfamily structureand attainment is explained by otherfamily back-ground variables, including smaller family size,being Catholic, higher levels of parental educa-tion, and being White. Men who are raised byboth natural parents are not advantaged educa-tionally, compared with those who grow up inother types offamily structures.A cohort analysisfor men that compares baby boomers with pre-baby boomers, however, suggests contradictoryeffects of family structure that deserve more ex-ploration.

Department f Sociology, 300 BrickerHall, The Ohio StateUniversity, 190 N. Oval Mall, Columbus, OH 43210([email protected]; arcel. @osu.edu).Key Words:divorce, earnings attainment,amily structure,gender, wo-parentfamilies.

How does family structure in childhood affect anindividual's subsequent socioeconomic well-being? We know that the absence of the fatherlowers cognitive test scores for young children(Mott, 1993) and thatbeing raised by a single par-ent may interfere with high school graduation(Coleman, 1988; McLanahan, 1985; Shaw, 1982)and with the availability of funding for higher ed-ucation (Steelman & Powell, 1991). McLanahanand Sandefur (1994) argue that children raised byonly one biological parent are deprived of eco-nomic and social resources, which negatively af-fects their future success. Boys who grow up inoriginal, two-parent families have a greaterchance of achieving a higher occupational statusthan those raised in families headed by women(Duncan & Duncan, 1969), and boys who areraised by their mothers have a greater chance ofending up in occupations with lower status thanthose their fathers held (Biblarz & Raftery, 1993).Amato and Keith (1991) indicate that childrenwho experienced parentaldivorce are more likelythan those raised in original, two-parent familiesto exhibit psychological, behavioral, social, andacademic problems. A great deal of evidence sug-gests that being raised without both natural par-ents involves costs for both children and the larg-er society. (See McLanahan & Booth, 1991, andSeltzer, 1994, for comprehensive reviews andDemo, 1992, for some counter arguments.)These studies do not tell us, however, whetherdivorce affects children when they mature to

Journalof Marriageand the Family 59 (May 1997): 419-433 419


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Journal of Marriage and the Familyadulthood. In addition, prior research on the ef-fects of family structure on socioeconomic out-comes frequently has failed to consider effects bygender by considering men only (Biblarz &Raftery, 1993; Duncan & Duncan, 1969; Green-berg & Wolf, 1982; Krein, 1986) or has beenlimited by confining attention to outcomes inyoung adulthood (Hill, Augustoniak, & Ponza,1987). This article extends previous work by pro-viding a clear causal picture of the effects ofchildhood family structure on adult socio-economic outcomes for both men and women.We differentiatetwo broad categories of fami-ly structure: those respondents who were raisedby both naturalparents most of the time until theywere 16 years old (original, two-parent familystatus) and those who report that they were not.We provide a rationale for using these broad cate-gories and expand on this definition. We study (a)how growing up with both naturalparents affectseducational attainment, occupational status, andearnings, compared with growing up in otherfamily structures, and (b) whether these effectsdiffer by gender.

THEORETICALUIDANCEHuman capital and status attainmentperspectivesprovide useful bases for studying the effects offamily structure on adult socioeconomic out-comes. Becker and Tomes (1986) propose thatparents influence the adult earnings of their chil-dren by "expenditures on their skills, health,learning, motivation, 'credentials,' and manyother characteristics" (p. S5). This theory positsthat the investments people make in educationand training will be reflected in returns that canbe measured in earnings (Becker, 1975). Invest-ments may be made on several levels, includingsocietal, parental, and individual levels. Familydisruption can diminish the investments that par-ents make in their children (Astone & McLana-han, 1991; Steelman & Powell, 1991).Two central explanations that suggest howfamily structure might have an impact on adultoutcomes are extensions of status attainment the-ories. Explanations of resource dilution lead us toanticipate that family disruption will have nega-tive consequences because disruption spreads thinwhatever resources the family has (Blake, 1989;Coleman, 1988). Families with less money, morechildren, or only one parent in residence mayhave fewer resources for support. In disruptedfamilies, not only is there normally a drop in in-come (Bianchi & McArthur, 1991; Hoffman &

Duncan, 1988; Holden & Smock, 1991; Peterson,1996; Weiss, 1984), but also time may be lackingfor activities, such as supervising children'shomework (Astone & McLanahan, 1991). Theseconditions also may apply to never-married,single-parent families. Explanations of resourcedilution do not offer insight into possible genderdifferences in the allocation of limited resourceswhen families are disruptedor when there is onlya single parent.Socialization explanations emphasize the im-portance of role modeling whereby parents fosteror deter attainment in their children through theirown examples (Hill et al., 1987; McLanahan &Bumpass, 1988). Because disruption often resultsin children being raised by their mothers, the lackof a male role model could affect boys negativelyby removing an example of how men achieve oc-cupational success. The lack of this model mayserve as a disincentive for boys to achieve.Through role modeling, girls may learn thatwomen are able to manage a family alone, whichmight lead to early or single childbearing. Social-ization models also predict that daughters couldrespond to the lack of a male figure in their livesby marrying earlier to fill a void (McLanahan &Bumpass, 1988). Early or single childbearing orearly marriages can mean that girls' educationsuffers. Additionally, the socialization modelspredict that disruption may weaken parents' so-cial control (Mueller & Pope, 1977). Althoughthe mechanisms for detrimental effects differ inhuman capital and key status attainmentexplana-tions, each theory predicts negative implicationsfor subsequent achievement.

Effects of Family Structure onEducational AttainmentWe know that the economic disadvantages expe-rienced by single-parent families can result inlower rates of high school graduationand dimin-ished parental funding of higher education. Addi-tionally, Astone and McLanahan (1991) foundthat children in stepfamilies and with single par-ents receive less parental encouragement and at-tention regardingeducational activities and have ahigher likelihood of leaving high school thanthose in original, two-parent families. They foundthat if socioeconomic status was controlled, therewere equally negative effects for single-parentfamilies and stepfamilies.McLanahan and Bumpass' (1988) study ofadult outcomes of women indicates that thosewho spend some time in one-parent families are

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Family Structure, Gender, and Earnings Attainment

more likely to marry and have children early, toget pregnant before marriage, and to divorce.Much of the effect of marital disruption onteenage motherhood is explained by its effect oneducation. Women whose families were disruptedduring their childhood are more likely to entermarriagesearlier;in these marriagesboth partnershave limited education (Mueller & Pope, 1977).McLanahan and Sandefur (1994) found that inWhite families, young men and women are morelikely to drop out of high school when they comefrom disrupted families, and women are muchless likely to complete college than men. They in-dicate that these findings are not consistent acrossall four surveys they study, however, and must beviewed as tentative. The effects are not as strongin Black and Hispanic families, and they reportno sex differences.

Short-term,detrimental effects also may havelong-term consequences. For example, Mott(1993) found that father absence predicts lowercognitive test scores and poorer emotional adjust-ment for 5- to 9-year-old children. Lower cogni-tive scores could lead to poor performance inschool with implications for future achievement.A number of people have studied how origi-nal, two-parent family status during childhood af-fects educational attainment. There are indica-tions that both men and women raised withoutboth naturalparents may experience some disad-vantages. Because education is an important partof our model and because earlier findings havebeen tentative, we include education to confirmand clarify priorwork.Effects of Family StructureonOccupational Status

In an early study, Duncan and Duncan (1969)found that for men, growing up in an original,two-parent family increases the probabilityof pur-suing an occupation that has relatively high so-cioeconomic status. More recently, Biblarz andRaftery(1993) found thatfamily disruptionaffectsoccupational mobility directly because men fromdisrupted backgrounds are more likely to enterlow-status occupations. According to Biblarz andRaftery, the intergenerationaltransmission of sta-tus and occupation takes place, in part, throughrole modeling, socialization, and sharing valuesbetween fathers and sons. When families breakup, the son's access to his father is diminished.Biblarz and Raftery imply that this lack of contactresults in the failure of sons to model the occupa-tional roles of their fathers. Coming from a dis-

rupted family both directly lowers men's occupa-tional destinations, compared with their father'soccupations, and weakens the association betweenthe father's occupation and his son's occupationaldestination, resulting in greater universalism.Comparableresearch on women is nonexistent, agap this article will fill.

Effects of Family Structure onEarnings AttainmentStudies of how family structureaffects adult earn-ings attainment have been limited. In research ona sample of 23-year-old men from the PanelStudy of Income Dynamics (PSID), Greenbergand Wolf (1982) found that parental marital dis-ruption caused reductions in the mother's hometime and family income, which resulted in lowerearnings for men in young adulthood. Effectswere strongest for Whites, but also were presentfor Blacks. Because only young men werestudied, this research may not give us a very goodindication of future earnings potential, nor does itaddress whether these effects are the same forwomen.

Using the Survey of Mature Women and theSurvey of Young Men to match mothers and sonsfrom the National Longitudinal Survey of LaborMarket Experience (NLS), Krein (1986) foundthat for a group of men who were 28-38 years oldin 1980, the earnings of those raised by a singleparent were negatively affected only indirectly-through lower educational achievement. Thosewho lived with one parent during preschool weremost likely to achieve less education by age 28.Krein cautions that additional negative effectsmay show up later because effects were measuredsoon after the completion of education.In a comprehensive study using the PanelStudy of Income Dynamics, Hill et al. (1987) ex-plored several aspects of attainment for a group of24-year-old men and women, including wages foryoung men and income-to-needs ratios for youngmen and women. They found that wages forWhite men were reduced because of net incomedeprivationin the family of origin caused by fam-ily disruption. Net income deprivation refers to asituation in which there is a net loss in family in-come that cannot be made up by the compensat-ing efforts of mother and children. The lower in-come constrains childhood developmentalactivities and ultimately reduces the respondents'adult attainments. Additionally, the familyincome-to-needs ratios for men and women indi-cated a reduction in the ability to meet family

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Journal of Marriage and the Familyneeds only for White and Black sons because ofnet income deprivation. There was no significantreduction on this dimension for the women in thestudy. Although this study included men andwomen, it looked only at income-to-needs ratios(not earnings) for women, and, again, it studied avery limited age range of adults early in theirworking years.

Alternative ExplanationsTo establish the impact of family structureduringchildhood on education, occupational status, andearnings, we consider alternativeexplanations forthe anticipated associations. Race is related toboth earnings and to original, two-parent familystatus. Blacks have higher rates of single parent-ing than Whites (Bumpass, 1984). Religion mayhave an impact on family status because ofRoman Catholic prohibitions against divorce.Smaller families may be better able to provide ed-ucational opportunitiesfor children (Blake, 1989;Downey, 1995). Married women work fewerhours and earn less than unmarried women (Pe-terson, R., 1989), and married men earn morethan unmarried men. These facts suggest thatmarital status is a possible alternative explana-tion. Finally, following human capital arguments,years of work experience and hours worked posi-tively affect earnings (e.g., Parcel, 1979; Parcel &Mueller, 1983), and thus are sensible controls.Regarding the potential for statistical interac-tion, we know that status attainment processesvary by gender (Parcel & Mueller, 1983). Womenobtain weaker returns to investments such as ex-perience or to characteristics such as occupationalstatus. Not only do adult socioeconomic out-comes vary by gender, but Morgan, Lye, andCondran (1988) report that families with malechildren are less likely to experience marital dis-ruption than families with only female children.Krein and Beller (1988) report that the negativeeffects of living in a single-parent family aregreaterfor boys than for girls. It is sensible, there-fore, to consider whether the effects of familystructure on several adult outcomes will also varyby gender.

HypothesesThe causal chain for studying earnings links fami-ly background characteristics to educational at-tainment, which leads to occupational status thathelps determine variations in earnings. Changes

in family background can have an impact on earn-ings, either directly or indirectly through their ef-fects on education or occupational status. Thesebasic effects should be in the same direction formen and women with one exception. Marriageshould have a negative effect on earnings forwomen and a positive effect for men.We expect educational achievement, occupa-tional status, and earnings attainmentfor childrenfrom original, two-parent families to be greaterthan for children from other family structures.Webelieve that the negative effects of disruptedfam-ily status on future earnings will differ for menand women. For women, the effects will operatethrough lower educational achievement. For menthe effects will operate through occupational out-comes.

METHODSSample and Measurement

We use data from the Panel Study of Income Dy-namics, a longitudinal survey that began in 1968.It consists of a representativesample of individu-als and the families in which they reside in theU.S. The study is conducted by the Survey Re-search Center (SRC) at the University of Michi-gan and consists of two independent samples withdifferent sampling frames. The first, known as theSRC sample, included nearly 3,000 families fromthe master sampling frame and was a cross-sec-tional, national sample using a stratified multi-stage sample of the civilian, noninstitutionalpop-ulation of the U.S. (Hill, 1992). The second sam-ple, from the Survey of Economic Opportunityand known as the SEO sample, sampled standardmetropolitan statistical areas and also includednonstandard metropolitan statistical areas in theSouth. It involved unequal selection probabilities(Hill, 1992). This sample included about 1,900families with low income in 1968 who were inter-viewed by the Census Bureau for the Office ofEconomic Opportunity (Inter-university Consor-tium for Political and Social Research, 1993). To-gether, both samples included a total of 4,802families in 1968. The members of the originalstudy have been followed through the years asnew families are formed when members leavetheir families of origin, resulting in over 7,000families in the 1989 interviews.Each year since the study began, data on in-come, family structure, and socioeconomic back-ground have been collected. If a married or per-

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manently cohabiting couple is in the survey, theman is normally designated the head of the fami-ly. Single women with or without children alsoaredesignated family heads. Surveys are conduct-ed annuallywith the head of the family when pos-sible, although in couples the wife is interviewedin some instances.In this study, we use a subsample of respon-dents (n = 3,523) from the 1989 panel (WaveXXII) who were family heads or wives in thatyear, who were aged 30-59, were from the civil-ian work force, and who reported an occupationand had earnings greater than zero in 1988. Thecombined PSID sample is a probability samplebut with unequal selection probabilities, makingweighting necessary in orderto representthe pop-ulation of the U.S. PSID weighting variablescompensate for both unequal selection probabili-ties and sample attrition(Hill, 1992). Because theoriginal families were selected in 1968, a recentincrease in Latino immigration is not reflected inthe primary sampling units, and that population isnot adequatelyrepresentedin this sample.We use only sample members from the origi-nal families in the study. Some were heads andwives in 1968, and others were children. Addi-tionally, although limited data are available on allother family members, the data for heads andwives are much richer and include detailed re-ports on family background, occupation, andearnings that are not available for other familymembers. Limiting the study to heads and wiveseliminates 156 otherfamily members who are pri-marily adult children still living with a parent orparents.We have limited the study to sample memberswho were 30-59 years old in 1989 because ourinterest is in people in their primary workingyears. By age 30, many individuals have complet-ed most of their education and have had time toestablish their own households or marry, thus be-coming heads or wives. Placing an upper agelimit of 59 on the sample helps assure that mostrespondents have not yet retired from the laborforce.We look primarily at those with earningsgreater than zero for two reasons. Methodologi-cally, we eliminate zero-earnersbased on an anal-ysis by Hauser (1980), who demonstrates that in-cluding values of zero on the log of earningsgives a distribution an extreme, negative skew.Substantively, we are interested in individualearnings of employed men and women. Previouswork has examined total family income, which

confounds marital status with an individual's owneconomic fate, especially for women. Althoughdata indicate that women frequently recover fromthe economic losses of divorce by remarrying(Duncan & Hoffman, 1985), there is some evi-dence to suggest that they are less likely to remar-ry if they are economically independent (Ambert,1983). In order to understandwhat (if any) biasesthis decision may introduce, we include a Heck-man correction to determine how family structureaffects the hazard of being excluded from thesample because of nonearner status ( Berk, 1983;Heckman, 1980). The results are reported sepa-rately near the end of this article.

VariablesOriginal, two-parent family status. Because ourrespondents were born between 1930 and 1959,detailed longitudinal data on their family historiesare unavailable. We must rely on their adult rec-ollections of their childhood. As Corcoran, Gor-don, Laren, and Solon (1991) note, retrospectiveaccounting can be inaccurate and thus injectssome error into the measurement of variables.This affects our measure of family status. Re-spondents were asked, "Were you living withboth of your naturalparents most of the time untilyou were 16?" Although this measure lacks detailabout durationor the ages at which children livedwith a single parent, there is some evidence thatthis may not matter. McLanahan and Sandefur(1994) found that the timing and duration of sin-gle parenting were unimportantfor understandinghigh school dropout rates or teen birth rates. Webelieve our measure succeeds in capturing theessence of the adult's perception of the familystructurein which he or she was raised. We be-lieve people know if their two natural parentswere together during their childhoods, and wethink this measure validly taps childhood familystructures.Adult children who were not raised in original,two-parent families may have a variety of experi-ences, including being raised by single, never-marriedmothers or by one parent after divorce orafter the death of the other parent. Or they mayhave been raised in stepfamilies or have grown upwith people other than their parents, such as fos-ter parents or grandparents. Although severalstudies have indicated that it may be importanttolook at the family structure of the disruptedfami-ly in which the child is raised (Acock & Kiecolt,1989; Biblarz & Raftery, 1993; Demo & Acock,

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Journal of Marriage and the FamilyTABLE 1. MEANS AND STANDARD DEVIATIONS OFORIGINAL, TWO-PARENT FAMILIES AND

OTHER FAMILY STRUCTURES FOR WOMEN AND MENWomen Men

Original,Two- OtherFamily OriginalTwo- OtherFamilyParentFamilies Structures ParentFamilies StructuresM (SD) M (SD) M (SD) M (SD)

FamilybackgroundNumberof siblings 3.58 (2.80) 3.62 (2.95) 3.50 (2.64) 3.50 (2.95)Catholicab .27 (.44) .20 (.40) .27 (.44) .18 (.38)Parents'educationbd 3.70 (1.56) 3.63 (1.50) 3.75 (1.52) 3.52 (1.28)Whiteabc .87 (.33) .72 (.45) .93 (.26) .80 (.40)Age 41.73 (8.05) 42.46 (8.43) 41.69 (8.13) 42.00 (7.97)EducationTotalyearsabc 13.34 (2.32) 12.75 (2.23) 13.66 (2.45) 13.38 (2.49)Dummies in years

0-1la .10 (.30) .17 (.37) .11 (.31) .14 (.34)12C .41 (.49) .43 (.50) .33 (.47) .36 (.48)13-15 .23 (.42) .25 (.43) .23 (.42) .21 (.41)16ac .12 (.33) .08 (.27) .18 (.39) .16 (.36)17+ac .14 (.34) .08 (.27) .16 (.36) .13 (.34)Marriedabc .74 (.44) .68 (.47) .85 (.35) .83 (.38)Occupational tatusa 42.86 (18.97) 38.58 (18.57) 41.82 (21.29) 40.92 (21.16)Yearsworkedc 14.37 (8.18) 15.42 (8.83) 20.99 (8.90) 21.62 (8.81)Hours worked n 1988C 1727.69 (672.43) 1782.02 (678.85) 2249.13 (645.35) 2194.38 (604.54)Hoursworkedperweekc 33.19 (12.93) 34.27 (13.05) 43.25 (12.41) 42.40 (11.63)EarningsLogged earningsabc 9.45 (.97) 9.34 (1.03) 10.21 (.79) 10.12 (.80)Dollars(in thousands)ac 17.91 (13.52) 16.03 (10.37) 36.18 (44.80) 32.48 (33.79)

(n = 1355) (n = 339) (n= 1497) (n = 332)aStatistically ignificantdifference between womenfromoriginal, two-parent amilies andotherfamilystructures.bStatistically ignificantdifferencebetween men fromoriginal,two-parent amilies andotherfamily structures.cStatistically ignificantdifference betweenmen andwomen.dParents' ducation is categorized from 1 to 8. 1 = 0-5 years of schooling, 4 = completion of high school, 8 = post-baccalaureatework.

1991; Kobrin & Waite, 1984; Wojtkiewicz,1992), a recent, detailed analysis of high schoolgraduationby Wojtkiewicz (1993) concludes thatconcern about family type overcomplicates theissue because major effects are observed with adichotomous indicator. In this study, we classifyall men and women who lived with both naturalparentsmost of the time until age 16 as from orig-inal, two-parent families. Those who state thatthey did not live with both natural parents untilage 16 are classified as from a perceived disrupt-ed family.Otherfamily background variables. Sex is coded1 if the respondent is a female; males are coded 0.Religion is coded 1 if Roman Catholic, 0 other-wise. Race is coded 1 if White, 0 otherwise. Mar-ried respondents are coded 1, 0 otherwise. Thenumber of siblings is the actual number of sib-lings during childhood as reported by the respon-dent. Parents' education is used to control for

family socioeconomic background.This is calcu-lated as the mother's education plus the father'seducation, divided by two. Although parental ed-ucation may not capture all aspects of the socio-economic status of the childhood family, general-ly it is highly correlated with occupational statusand takes advantageof available information.Education and occupational status. We use totalyears of education as a dependent variable. As anindependent variable, we create categories to rep-resent credential completion, with categories for0-11 years, 12 years (the comparison group),13-15 years, 16 years, and 17 or more years.Dummy variables allow us to examine relativedifferences in the importance of levels of educa-tion for men and women on occupational statusand earnings. Occupational status is based onStevens and Featherman's (1981) MSE12 revisedindex of Duncan's original, 1950, (SEI) index ofoccupational socioeconomic status.

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Earnings. Earnings are logged to reduce the im-pact of outliers and to allow interpretationof rela-tive differences, rather than absolute amounts(Peterson, T., 1989). Peterson reviews the myriadof functional forms that sociologists have used toestimate earnings equations, including log linearforms. He mathematically demonstrates that con-trolling for annual hours worked on earnings isequivalent to setting an hourly wage, which con-founds the exploration of differences in earningsbetween subgroups.We agree with his arguments,and we control for hours worked by setting thecoefficient to 1 for that variable. Peterson alsonotes thatearnings inequality is generatedboth bydifferences in hourly wages and by differences inhours worked. When logged earnings is the de-pendent variable and we control for hoursworked, once we fix hours worked, all variation isdue to the variation in the hourly wage rate. Thisallows the dependent variable-earnings, to beinterpreted in terms of the hourly wage, and theindependent variables then are interpreted interms of the hourly wage, a convenient interpreta-tion. If earnings is the dependent variable and ifwe simply control for hours worked, it impliesthat the average hourly wage rate depends on thenumbers of hours worked when it actually de-pends on other factors, such as education and ex-perience. (See T. Peterson, 1989, for the deriva-tion of equations.) This variable, adultearnings, isbased on Parcel (1979), and includes "earningswhen employed by others, income from self-employment, and other income derived from re-spondents' own labors" (p. 267). The number ofyears worked since age 18 includes full-timework that the respondent reported.

Missing values. We use mean substitution formissing values. The percentage of missing valuesranged from 0 for number of siblings, marital sta-tus, occupational status, hours worked, and earn-ings to 7.5% for parent education. Most variableswere missing values in fewer than 1%of cases.

Analytic StrategyAfter presenting basic descriptive data, the centralanalytic strategy is a series of multiple regressionanalyses where variables are entered in the orderdetermined by theory. Based on this fully recur-sive model, we perform separate analyses for menand women. We then consider interactions and in-vestigate possible differences between earnersand nonearners in a hazardanalysis.

DESCRIPTIVEESULTSMeans and StandardDeviations

We present means, standarddeviations, and nota-tions of differences of means tests in Table 1.Both men and women who were raised in origi-nal, two-parent families complete more totalyears of education and have higher logged earn-ings than those from other family structures.Some of the differences are modest. Additionally,a comparison of means for women and men indi-cates that men achieve more years of education athigher levels than women, and they have higherearnings. Men and women are similar in levels ofoccupational status. Bivariate correlations be-tween the respective dependent variables and pre-dictors are in the expected directions and general-ly statistically significant (data available from thefirst author).

TABLE 2. YEARS OF EDUCATION REGRESSED ON FAMILY STRUCTURE(EQUATION 1) ANDFAMILY BACKGROUND (EQUATION 2) FOR WOMEN AND MENWomen Men

Equation1 Equation2 Equation1 Equation2b P b P b P b PIndependentVariables (SE) (SE) (SE) (SE)

Original, wo-parent amily .59** .10 .57** .10 .28t .04 .10 .02(.14) (.13) (.15) (.13)Constant 12.75 10.98 13.38 11.83R2 .01 .22 .00 .21AdjustedR2 .01 .22 .00 .21

Note: Forwomen,n = 1,694. Formen,n = 1,829. InEquation1 original, two-parent amily status s enteredalone. Equa-tion 2 includes original,two-parent amily status and otherfamily backgroundcharacteristics, ncluding logged numberofsiblings,Catholic, ogged parental ducation,White.tp < .05. *p < .025. **p < .005. One-tailed est.

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TABLE. OCCUPATIONALTATUS EGRESSEDNFAMILYTRUCTUREEQUATION),FAMILYACKGROU(AND EDUCATION (EQUATION 3) FOR WOMEN AND MENWomen

Equation1 Equation2 Equation3 Equation1b P b Pb 3 b 3IndependentVariables (SE) (SE) (SE) (SE)

Original, wo-parent amily 4.31** .09 3.39** .07 .71 .01 .94 .02(1.15) (1.08) (.92) (1.28)Numberof siblings -.70** -.10 -.30* -.04(.16) (.14)Catholic 2.05* .05 3.14** .07(1.00) (.84)Parents'education 3.68** .30 .82** .07(.29) (.27)White 3.43** .07 3.50** .07

(1.28) (1.08)Educationa0-11 years -10.48** -.18(1.23)12years(omittedcategory)13-15 years 6.20** .14(.95)16 years 17.44** .30(1.25)17+years 26.76** .47(1.23)Constant 38.56 24.83 30.17 40.89R2 .01 .14 .40 .00AdjustedR2 .01 .14 .39 -.00

Note: Forwomen,n = 1,694. Formen,n = 1,829.Educationdummies arestatistically ignificantas agroup.Women,F = 174.83, significantF .0000. Men, F = 279.23, significantF .000*p< .025. **p < .005. One-tailed est.


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Family Structure, Gender, and Earnings AttainmentTABLE . LOGGED ARNINGS EGRESSEDNORIGINAL,WO-PARENTAMILY TATUSEQUATION), FAMILYBACKGROUND(EQUATION), EDUCATIONEQUATION), MARITAL TATUSEQUATION), ANDWORK EQUATION) FORWOMEN

Equation1 Equation2 Equation3 Equation4 Equation5Independent b t ratioa b t ratio b t ratio b t ratio b t ratioVariables (SE) (SE) (SE) (SE) (SE)Original, wo-parent .11' 1.84 .1ll 1.77 .03 .45 .03 .55 .03 .89family (.06) (.06) (.06) (.06) (.04)Logged numberof -.04* -2.19 -.01 -.85 -.01 -.73 -.02t -1.92siblings (.02) (.02) (.02) (.01)Catholic .09t 1.66 .12* 2.31 .12* 2.36 .08* 2.21(.06) (.05) (.05) (.04)Loggedparents' .29** 5.19 .04 .68 .02 .42 .04 1.06education (.06) (.06) (.06) (.04)White .03 .41 .01 .08 .10 1.52 .07 1.51(.07) (.07) (.07) (.05)Educationb

0-11 years -.49** -6.29 -.52** -6.74 -.15** -2.74(.08) (.08) (.05)12 years (omitted)category)13-15 years .12* 1.96 .12* 1.97 .07t 1.71(.06) (.06) (.04)16 years .35** 4.14 .33** 4.19 .27** 4.99(.08) (.08) (.05)17+ years .60** 7.71 .59** 7.73 .36** 6.43(.08) (.08) (.06)Married -.33** -6.37 -.11** -3.10(.05) (.04)Logged occupational .37** 9.68status (.04)Logged yearsworked .27** 8.39since 18 (.03)Loggedhoursworked 1.00 fixed(.00) parameterConstant 9.34 8.97 9.25 9.43 -.15R2 .00 .03 .09 .12 .61AdjustedR2 .00 .02 .09 .11 .60

Note:n = 1,694 women.atratios areshowninsteadof standardized s.bEducation ummiesarestatisticallysignificantas agroup,F = 31.63, significantF .0000.tp .05. *p


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Journal of Marriage and the FamilyTABLE5. LOGGED ARNINGS EGRESSED NORIGINAL,WO-PARENTAMILY TATUSEQUATION), FAMILY ACKGROUND(EQUATION), EDUCATIONEQUATION), MARITAL TATUSEQUATION), ANDWORK EQUATION) FORMEN

Equation1 Equation2 Equation3 Equation4 Equation5Independent b t ratioa b t ratio b t ratio b t ratio b t ratioVariables (SE) (SE) (SE) (SE) (SE)Original, wo-parent .08t 1.76 .02 .44 -.00 -.01 -.00 -.06 -.03 -.69family (.05) (.05) (.05) (.04) (.04)Loggednumberof -.05** -3.90 -.03* -2.14 -.02* -2.00 -.02* -1.97siblings (.01) (.01) (.01) (.01)Catholic .32** 7.65 .30** 7.52 .29** 7.36 .23** 6.43(.04) (.04) (.04) (.04)Loggedparents' .31** 7.18 .09* 2.08 .11* 2.54 .07t 1.78education (.04) (.05) (.04) (.04)White .24** 3.87 .22** 3.59 .17** 2.82 .09t 1.70(.06) (.06) (.06) (.05)Educationb

0-11 years -.23** -3.79 -.22** -3.64 -.20** -3.72(.06) (.06) (.05)12years(omittedcategory)13-15 years .16** 3.45 .17** 3.69 .04 1.02(.05) (.05) (.04)16years .43** 8.39 .43** 8.44 .20** 3.87(.05) (.05) (.05)17+years .53** 9.55 .54** 9.95 .18** 3.12(.06) (.05) (.06)Married .29** 6.06 .11* 2.51(.05) (.04)Logged occupational .33** 8.60status (.04)Loggedyearsworked .20** 4.46since 18 (.05)Loggedhours worked 1.00 fixed(.00) parameterConstant 10.12 9.53 9.66 9.43 .38R2 .00 .08 .16 .18 .33AdjustedR2 .00 .08 .16 .17 .32

Note:n = 1,829men.atratiosare showninstead of standardized3s.bEducation ummies arestatisticallysignificantas a group,F = 41.51, significantF .0000.tp


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effects of graduate education are minimal formen. Men receive relatively more benefits from a4-year college degree than for higher levels of ed-ucation. Women receive relatively more returnsfrom graduate work. Being married works tomen's advantage as far as earnings are concerned.So do higher occupational status and years ofwork experience. For theoretical reasons men-tioned earlier, we fixed the coefficient for hoursworked to 1 for the earnings analyses. Althoughthe coefficients for effects more clearly predictwage attainment in our model than in an unre-strictedmodel, significance of predictors is virtu-ally identical in both models.

AdditionalAnalysesInteractions. Growing up in a home without bothnaturalparents appearsto disruptwomen's educa-tional attainment but not men's. In order to deter-mine if there is a significant interaction betweengender and two-parent family status for educa-tional attainment,we combined men and womenin the sample and created a multiplicative genderx two-parent family interaction variable. Whenadded to the equation, the interaction is signifi-cant at p < .05, two-tailed test. Clearly, educationis more disrupted for women who did not livewith both parents during childhood than for men.There is no significant interaction of two-parentfamily status and gender on occupational status orearnings. In addition, we created a multiplicativeterm composed of race and two-parent family sta-

tus. This term was not significant for any of thedependent variables.The finding of no effects of growing up in anoriginal, two-parent family on educationalachievement, occupational status, or earnings at-tainmentfor men is provocative because both the-ory and prior empirical work indicate that livingin two-parent families during childhood shouldhave positive implications for adult outcomes. Inresponse to these findings, we divided our sampleof men into two groups: those born prior to thebaby boom (before 1946) and who entered a rela-tively friendly labor market, and baby boomers,born from 1946 through 1959, who entered atighter labor market. This analysis suggests thattwo-parent family status had negative effects onearnings for men in the older group and positiveeffects for the younger men (see Table 6).It is possible that in the whole group of men,these effects cancel each other out, resulting inthe finding that two-parent family status has nosignificant effect on outcomes for men. Althoughthere are opposite effects on earnings for the twogroups of men, for both groups the effect is ex-plained when occupational status, years worked,and hours worked are added to the model. Effectsdo not occur through education, as is the case forwomen. These results suggest that the finding ofno significant effect of two-parent family statusduring childhood for men as a whole should bereexamined. A similar analysis by cohort for thewomen (not shown) indicated the effect of family

TABLE 6. COMPARISONOF YOUNGER MEN AND OLDER MEN, LOGGED EARNINGS REGRESSED ONORIGINAL, TWO-PARENT FAMILY STATUS (ABBREVIATED EQUATIONS)Equation1 Equation2 Equation3 Equation4 Equation5

Equation1 andOriginal,Two- Family Background Equation2 and Equation3 and Equation4 andParentFamily Characteristics Education MaritalStatus Work Variablesb b b b b(SE) (SE) (SE) (SE) (SE)

Youngermen (n = 1,114) .24** .12* .10t .09t .09(.06) (.06) (.06) (.06) (.05)Older men (n = 715) -.16* -.13t -.16* -.16* -.08(.08) (.08) (.07) (.07) (.06)All men (reference) .08t .02 -.00 -.00 -.03(.05) (.05) (.04) (.04) (.04)

Note: Equation1:Original,two-parent amily status s entered alone. Equation2: Original,two-parent amily status andother family backgroundcharacteristics, ncluding logged numberof siblings, Catholic,logged parents'education,White.Equation3: Original,two-parent amily, otherfamily background haracteristics,andeducation.Equation4: Original,two-parentfamily, other family backgroundcharacteristics,education, and married.Equation5: Original, two-parent family,other family backgroundcharacteristics, education, married,and work variables, including logged occupational status,logged yearsworked, ogged hoursworked(set to 1).tp .05. *p


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Journal of Marriage and the Familystructure during childhood on attainment out-comes was similar for both groups.Earners versus nonearners. One might argue thata sample including only those with earnings whodeclare occupations would fail to reveal the entireeffect of original, two-parent family status duringchildhood on adult outcomes. Suppose, for exam-ple, that children raised in single-parent familiesare less likely to become earners. To determinewhether the effects of growing up in a two-parentfamily differ for those who are excluded from thesample because of their status as nonearners, weconduct a hazard analysis. We construct a vari-able to represent the hazard of being excludedfrom the analysis because of nonearner status,given that respondents who are included are earn-ers. The hazardof nonemployment measures a la-tent trait that varies among those who are em-ployed. Leaving it out of the model could biassome of the observed effects. In this case, thehazard of nonemployment is constructed from avariable indicating the respondent's assessment ofpersonal health because there is a positive corre-lation between health and earnings. We believethat the effect occurs simply because poor healthaffects the chance of being excluded from thesample. We first add all nonearning heads andwives back into the sample (n = 4,576) and per-form a probitanalysis that predicts the probabilityof being in or out of the sample because of earnerstatus. This equation includes key independentvariables as well as the indicatorfor health. Theseresults are used to compute the hazard of nonem-ployment. We then run the regressions for earn-ers, including the hazardvariable. This analysis isperformed with Limdep 6.0, which does not per-mit the restriction of the coefficient of hoursworked to 1, so we have estimated another modelfor comparability (details available from the firstauthor).Our results indicate that the hazard of beingexcluded differs for women and men. The hazardis insignificant for women, indicating that limit-ing our sample to women who are earners doesnot bias our estimates. For men, the hazard ofbeing a nonearner is significant, but the signifi-cance levels of most of the independent variablesremain stable. The equation controlling for sam-ple selection bias reveals that the positive effectsof parents' education, being married, and beingWhite in the original analysis are due to sampleselection. In addition, once we control for thehazard of nonemployment, low levels of educa-

tion no longer have a statistically negative effecton earnings for men. In sum, respondents fromhigher socioeconomic backgrounds, those whoare White or married are more likely to be earn-ers, hardly a surprising finding.Because some of the differences in familystructure act through education, we included allnonearners with earnersin a regression on educa-tion for men and women. The results were essen-tially the same as in the main analyses. For exam-ple, in the main analysis, women from original,two-parent families had significantly higher lev-els of education than those from other familyforms, even when other family background char-acteristics were controlled. That was not the casefor men. For men, the effect of two-parent familystatus on education was explained by other familybackground variables. When nonearners were in-cluded in the analysis, the same pattern appeared.

DISCUSSIONOur interest was to explore the effects of growingup in an original, two-parent family on the earn-ings attainmentprocess and to explore specifical-ly how this process differs by gender. We predict-ed that men and women from two-parent familieswould fare better than those from other familystructures. This is supported. Further, we foundthat there were gender differences in the effects offamily structure during childhood on adult out-comes. When all earners between the ages of 30and 59 in 1989 are considered, growing up in atwo-parent family is more important for womenthan for men. Women's education suffers whenthey are raised in disrupted families, and thosedisadvantages carry through to occupational andearnings outcomes. The same effect is not presentfor men. The positive association between two-parent family status and education for men is ex-plained by other family background variables. Inaddition, when the men are divided into two co-hort groups, growing up in a two-parent familyhas positive implications for the earnings of themen born during the baby boom, and it has nega-tive effects on that outcome for men born prior tothe baby boom. Although our data set does notallow the investigation of the cause of disruption,we know that most family disruptionfor the oldergroup was caused by death rather than divorce.We speculate thatbecause the pre-baby boom andbaby boom cohorts entered the labor marketunder dramaticallydifferent economic conditions,the interplay of family structure and economic

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opportunity when the labor market was enteredmay have had subsequent effects on their earn-ings attainment.In conclusion, these findings support status at-tainment and human capital theories for thewomen in the study. Two-parent family status canbe viewed as a valuable family background char-acteristic that enhances educational opportunitiesfor daughtersand has subsequent implications foroccupational status and earnings. The value oftwo-parent family status as a resource for men ismuch less certain. Overall, for the men in the

study, being raised in a two-parent family offersno apparent benefit for earnings attainment out-comes. However, the results of the cohort studysuggest that the state of the labor market may ex-plain differences in the effect of family structureon the results for men. For the younger men, thecohort analysis supports earlier findings ofBiblarz and Raftery (1993) that men whose fami-lies are disrupted by divorce are disadvantagedintheir occupational endeavors. What is clear fromthese data, however, is that men's education is notcompromised by disrupted family status duringchildhood the way women's education is. The co-hort analysis also suggests that the positive andnegative effects of various family structurescouldbe historically specific. Furtherresearch is neededon social supports available to disruptedfamiliesof earlier times.It is useful to consider what implications thesefindings might have for successive cohorts ofyouths who experience divorce during childhood.The divorce rate is likely to remain quite high inthe near future (Martin & Bumpass, 1989), andgreat numbers of children will spend part of theirchildhood in single-parent families (Bumpass &Rindfuss, 1979; Furstenberg & Nord, 1985). Ifthe processes we have modeled here hold foryounger cohorts, women from disrupted familiesmay be more at risk for attaining fewer years ofschooling than are men from disrupted families.The specific mechanisms at work could includewomen not receiving the financial support thatmen do to attainadditional schooling. In addition,women may not receive the parental encourage-ment and social supportneeded to aspire to higherlevels of education or to complete degrees. Giventhe strong impact of schooling on earnings, lowerlevels of educational attainment have clear impli-cations for financial well-being. Although girlsnot raised in original, two-parent families lose, onaverage, only slightly more than a half year of ed-ucation, as Krein (1986) points out, which half a

year is lost is important. If that time makes thedifference between completing and not complet-ing high school, for example, it could make a bigdifference in later socioeconomic outcomes.For these reasons, parents and policymakersneed to consider the possibility that girls fromdisruptedfamilies may be especially vulnerable tocompromising educational aspirations and attain-ment. This possibility suggests the need for par-ents to explicitly consider how marital dissolutionmight affect the educational attainment of theirchildren, particularly their daughters. Theseyouths especially may need encouragement fromparents and schools to pursue higher levels of ed-ucation.

Many of today's children will be raised insingle-parent and blended families, and it is vitalthat we have as clear an understanding of theshort- and long-range implications of thesechanges as possible. This study alerts us to thelong-range implications of family structure andreminds us that the mechanisms and the impactcan differ for women and men. We also must re-main cognizant of the difficulties families andchildren face when there is conflict or lack of re-sources that may bring disadvantages to thosechildren concerned.

NOTEAn earlierversionof this articlewas presentedat the1995annualmeetingof theAmericanSociologicalAs-sociation n Washington,D.C. We thankRobertKauf-man,ElizabethG. Menaghan, ohnMirowsky,Cather-ine Ross, and two anonymous eviewers or commentsandadviceonearlierdrafts.

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Ed 1000 Attainment in Social Edu

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