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The Impact of Child Care Subsidies on Single Mothers’Work Effort

April CrawfordUniversity of Houston

Abstract

Access to affordable child care is a formidable barrier to employment for many families. Following passageof the Personal Responsibility and Work Opportunity Reconciliation Act (PRWORA) in 1996, fundingfor child care assistance increased dramatically. Using data from the National Survey of America’s Fam-ilies (NSAF), the impact of child care subsidies on single mother’s work effort is evaluated. The resultsindicate that not only are women more likely to be employed when they receive child care subsidies, theyare also more likely to be employed full time. Single mothers who reported receiving assistance with childcare worked, on average, 9.4 hours more per week than mother’s who did not receive help. While growthin funding for child care has slowed in recent years, findings like the ones presented here highlight theimportance of affordable child care in the employment decisions of mothers with young children.

In the wake of the 1996 Personal Responsibility and Work Opportunity Reconcilia-tion Act (PRWORA), states have had to increase the amount of money they spendon child care to support the tougher work requirements ushered in by the reform.While the nation saw a dramatic increase in child care spending in the early yearsfollowing reform, spending has hit a plateau, and further growth is not expected(Rodgers, 2006, p. 184). Access to affordable child care is a critical precursor toemployment for many parents. Yet there is considerable variation in support amongstates, in terms of subsidy size, copayments for care as a percentage of a recipient’sincome, and in the percentage of low-income children served across states. Highchild care costs coupled with low wages can reduce the value of employment formany low-income families, particularly for single parents (Ford, 1996). In fact, poorworking families typically spend a third of their incomes on child care, whileworking class and middle-income families spend as little as 13% and 6%, respectively(Hofferth, 1999). Studies have clearly shown the negative relation between mater-nal employment and child care costs (Kimmel, 1998; Mason & Kuhlthau, 1992).High child care costs have been demonstrated to keep some women from joiningthe workforce entirely while causing others to leave their jobs prematurely (Blau &Robins, 1989). A study of Aid to Families with Dependent Children (AFDC) recipi-ents, carried out by Siegel and Loman (1991), indicates that many women who haveproblems with child care return to public assistance or quit education and trainingprograms as well. This financial disincentive to work is likely compounded by thepoor quality of child care available to parents with less money to spend.

Using data from the National Survey of America’s Families and reports on ChildCare Development Fund spending, the impact of child care subsidies on low-income families work effort is examined. While the data presented includes bothfamilies with two adults at home and single parents, the balance of the articlefocuses on single mothers with children under 5. While growth in employmentamong this group has been considerable in recent years, this group of women isstill among the least likely to be employed and, arguably, the most in need of childcare assistance.

Review of Policy Research, Volume 23, Number 3 (2006)© 2006 by The Policy Studies Organization. All rights reserved.

Background

National Changes

Transforming welfare into a supported work program requires that recipientparents have access to affordable child care. As shown in Figure 1 (Childcare andDevelopment Fund [CCDF], 2003, p. 4), overall spending on child care hasincreased dramatically since the 1996 reform. Under PRWORA, four child careprograms, AFDC/JOBS Childcare, Transitional Childcare, and At-Risk Childcare,were consolidated into the Childcare and Development Fund (CCDF), which isadministered by the Childcare Bureau under the Administration for Children andFamilies.

Flexibility allowing states to transfer up to 30% of their Temporary Assistance toNeedy Families (TANF) dollars into the CCDF, as well as to spend TANF fundsdirectly on child care has also contributed to the dramatic increase in overall childcare spending. In 2001, 42 states transferred funds from TANF to CCDF, sometransferring more than 100 million dollars (CCDF, 2003, p. A-3). States also spent1.6 billion TANF dollars on direct child care assistance in 2001. The substantialincrease in child care funding has impacted the number of children served, thesize of parent copayments, and the quality of child care in many states (Rodgers,2006, p. 183). These changes should help smooth the transition to work for manywelfare recipients and assist working families living at the margins of poverty.

State-Level Variation

The PRWORA allows states considerable control over spending on child care inorder to meet the unique needs of each state’s diverse population. This flexibility

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Figure 1. CCDF- and TANF-related Child Care Funding

has resulted in significant variation in spending patterns across the states. As seenin Table 1, states vary considerably on per child spending, the size of requiredparent contributions to care, and in the percentage of the low-income populationin a state served. States are permitted to adjust their eligibility requirements towhatever level they deem necessary, provided they serve families at or below 85%of the state median income. It appears that most states, in order to serve parentsmost in need, target families with earnings far lower than the federal maximum.Prior work by Meyers, Heintze, and Wolf (2002) indicates that current welfarerecipients, particularly those enrolled in JOBS programs, are the individuals mostlikely to receive child care subsidies, though subsidization even among this groupis still a “rare event” (p. 165). Targeting this group for child care assistance meansthat many working families at or near the poverty line are not receiving the assis-tance they may need to stay employed and off welfare. On average in 2002, stateswere providing child care assistance for only 14% of those children who were eli-gible (CCDF, 2003). It appears that a lack of information about child care subsidiesamong those eligible and the complexity of the process for receiving assistance contribute to low rates of subsidization (Meyers & Heintze, 1999). There is also evidence to suggest that should these barriers be removed there would still be a considerable shortfall of coverage unless funding was also increased (Long &Clark, 1997).

Sample

The data used to test the impact of subsidy receipt on employment in this studywas collected during the third wave of the National Survey of America’s Families(NSAF), which has been administered to over 100,000 respondents since 1996. Thethird round of the survey was conducted in 2002, and contains information on40,000 families. Households were sampled using both random digit dial for tele-phone interviews, and census block data to obtain a random selection of dwellingsto conduct in-person interviews. Census block data, and initial screening probes onincome level, allowed the Urban Institute to heavily oversample low-income house-holds, making the sample a rich source of information about social program use.While individuals in all 50 states and the District of Columbia were sampled,Alabama, California, Colorado, Florida, Massachusetts, Michigan, Minnesota, Mis-sissippi, New Jersey, New York, Texas, Washington, and Wisconsin were designatedas focal states. More than half the country’s population resides in these states, which

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Table 1. State Child Care Variation 2002

Low High Average

Spending per child 2,667 12,480 5,726South Carolina Utah

Copay as percent of income Less than 1% 22% 8.27%New Hampshire Oregon

Percent children served 6% 53% 14%with working parents Mississippi Delaware(<62%SMI*)

*SMI-State median income.

represent a broad range of state-level characteristics, approaches to governance,and social policies (National Survey of America’s Families, 2002). The survey cap-tures critical aspects of the economic, health, and social characteristics of childrenand adults under age 65. The initial wave of the survey was conducted in 1996,concurrent with PRWORA’s inception. It utilizes a comprehensive battery of itemsconcerning the use of a variety of programs including AFDC/TANF, Medicaid, andUnemployment. Detailed information collected included prior program involve-ment, current use, and the reasons why those eligible are not currently receivingservices.

The current analysis is concerned with the impact of child care subsidies onemployment, particularly among single mothers. Models are initially estimated onall sampled parents who have children under the age of 5, regardless of marital orcohabiting status. There is, however, considerable variation in eligibility require-ments between states that could limit subsidy qualification for two-parent house-holds. In light of this, the models are reestimated on a reduced sample consistingonly of single mothers. Single mothers are the group most likely to receive, andbenefit from, child care subsidies. When children enter the public school systemmaternal employment increases dramatically, therefore, parents whose children areall over the age of 6 have been excluded from this analysis.

Methods

Since the reform in 1996, empirical evidence has continued to suggest the impor-tance of child care in a parent’s decision to work. Blau and Tekin’s 2001 studyshows a modest subsidy impact on employment and more substantial effects if oneconsiders participation in education and training activities. Using binary measuresof work participation, they find that subsidy receipt among single mothers leads toa 5% increase in the likelihood of employment, and a 13% increase in the likeli-hood of being involved in a broad class of work activities. These findings, whilemodest, are encouraging.

The data analyzed by Blau and Tekin comes from the 1997 wave of the NSAF.While spending on child care had increased in the years leading up to the reform,the majority of the increase in spending occurred in the years after the 1997 datawas collected. Blau and Tekin note that the 1997 data was collected when subsidieswent almost entirely to welfare recipients or recent leavers of the welfare system.These individuals are among the most difficult to place in jobs. Since the PRWORA,states have been able to expand the population of those eligible to receive subsi-dies, so we should expect the impact of subsidies to look stronger now than it didin 1997.

To facilitate comparison with prior work on subsidy receipt, the first model esti-mated uses the same NSAF item Blau and Tekin used in their study, which is coded1 for persons currently working more than 15 hours per week, and 0 for all others.Since the current investigation is primarily interested in whether or not subsidieshave the ability to move mothers closer to independence, the same models wereestimated with a full-time work indicator coded as 1 for persons working more than35 hours per week and 0 for all others. While previous studies have tended toinclude enrollment in training and education activities, the dependent variables

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used in this study focus solely on employment. Next, the models are estimatedusing the reported number of hours a respondent has worked per week over thepast year. This higher level of data is generally preferred over the dichotomouswork variables, which ignore a considerable amount of variance present in thesample.1 Logit is applied when working with the part- or full-time employmentdummies, and ordinary least squares (OLS) is used to estimate the hours workedper week model.

The primary variable of interest is a simple binary indicator of subsidy receipt.2

The number of mothers in the sample with children under 5 receiving a child caresubsidy was 451, which represents nearly one-third of the sample. The sample ofsingle mothers range in age from 16 to 55 and the mean age for subsidy receipt is28 years. The basic hypotheses state that parents who receive a subsidy will be morelikely to work, to work more hours per week, and to work full time. Additionally,it is hypothesized that the value of child care subsidies will be greater for womenwho are not married or cohabiting than it will for other subsidy recipients. Themodels are estimated first on the sample of all parents of children under 5, andthen again for single mothers. This sample reduction resulted in the exclusion of54 men who received a subsidy, and 1,700 married or cohabiting respondents whomay or may not have received a subsidy.

Individual- and state-level control variables are applied. Age and gender areapplied as basic demographic controls. It is likely that work will be positively relatedto both age and being male. Age is a continuous variable and gender is coded as0 for males and 1 for females. Dummy variables indicating marital and cohabitingstatus were combined to generate a single status variable, where 0 is coded whena respondent is either married or cohabiting, and 1 represents single parents. Hereit is hypothesized that single parents will be more likely to work and work morethan married or cohabiting parents who have another potential earner in thehome. A final control variable specified indicates whether or not the respondentreceives TANF. This variable is coded 1 for yes, and 0 for no. As discussed above,prior work on the determinants of subsidy receipt show that current or recentwelfare recipients are often targeted for child care assistance. Although in thissample the correlation between TANF and child care subsidy indicates that fewrespondents are participating in these programs jointly, it should still be interest-ing to see what impact TANF receipt has on employment relative to child care sub-sidies. Variations on race and ethnicity were tried in earlier models but were alwaysnonsignificant, and as such are not included in the final models.

Several state-level indicators were included in the analysis, with little success, inan effort to control for contextual effects specific to individual states. Several eco-nomic indicators were tried, including state unemployment and state unemploy-ment for women, but only per-capita income was retained in the analyses presentedhere. Also included is a measure of welfare plan sophistication based on the gen-erosity and coverage of state plans. This variable, developed by Meyers, Gornick,and Peck (2001), ranges from 1 to 5, with 5 being the least sophisticated plan.Finally is a dichotomous variable developed by Soss that indicates whether or nota state conditions TANF aid on tougher work requirements than those set by thefederal government (Soss, Schram, Vartanian, & O’Brien, 2001). This variable iscoded 0 for no work requirement, and 1 if work is required. The inclusion of the

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sophistication and work requirement variables should increase our confidence thatsubsidies have an independent effect from other features of a state’s social programs.

Results

Logit Models and Predicted Probabilities

The results of the binary logit models are presented in Table 2, and the predictedprobabilities associated with these models are shown in Table 3. In all cases, theimportance of child care subsidies is strongly supported. Beginning with a com-parison of Blau and Tekin’s (2001) results on an earlier wave of NSAF data, we seethat the impact of child care subsidies on employment has grown since 1997. Blauand Tekin estimated that in 1997 a single mother was about 5% more likely to beemployed at least 15 hours per week if they received a child care subsidy. Usingthe same dependent variable and subsidy indicator on the 2002 data, we can esti-mate the predicted probability of working more than 15 hours per week as 0.75for single mothers receiving a subsidy, versus 0.54 for those not receiving a subsidy.3

When the model is reestimated as a probability of working 35 or more hours perweek we see the raw value of the subsidy decrease by 16 points, yet in this modelthe subsidy is still what seems to make full-time employment more likely than not.In the single mother sample, the predicted probabilities of working full time movefrom 0.38 when no subsidy is received to 0.53 when one receives a subsidy. All other individual-level control variables also have an effect that is statistically significant.

When the models are estimated on the full sample of parents, regardless ofgender and marital cohabiting status, the picture is very similar to the onedescribed above. The main differences being that men are far more likely to be

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Table 2. Logit Coefficients

Full Sample Single Mothers

Logit-Work Logit-Full Time Logit-Work Logit-Full Time

Age 0.20** 0.02** 0.03* 0.04**(0.005) (0.006) (0.008) (0.008)

Gender −1.94** −2.15** — —(0.151) (0.13)

Single 0.70** 0.77** — —(0.08) (0.08)

TANF −1.01** −1.03** −1.07** −1.07**(0.22) (0.26) (0.24) (0.28)

Child care subsidy 0.84** 0.55** 0.92** 0.62**(0.12) (0.12) (0.14) (0.14)

Per capita income — — — —Sophistication 0.018 0.06** −0.02 0.04

(0.02) (0.02) (0.03) (0.03)Work requirement 0.27** 0.09 0.03 −0.05

(0.08) (0.09) (0.12) (0.12)Constant 0.63 −0.042 −0.538 −1.64

(0.26) (0.25) (0.28) (0.29)Observations: 3,127 3,127 1,372 1,372

OLS/Logit coefficients, standard errors in parentheses; **p < 0.001; *p < 0.05.

employed than women, and single women are far more likely to be employed than women who are married or cohabiting. The probability of working part time or full time is 40 points higher for men than women. The predicted proba-bility of women working at least part time increases from 0.61 to 0.76 if they aresingle, and the probability of working full time increases from 0.44 to 0.62 whensingle.

The impact of age and TANF conform to expected patterns; as respondents agethe probability of employment increases. Age effects are particularly pronouncedamong single mothers with children under 5. In this group the probability ofworking increases by about 10 points with every 10 years of aging. Interestingly,the probability of a single mother with children under 5 working full time does noteven reach 0.50 unless the woman is over 35. Single mothers, without adequatechild support, who are not working full time are likely living in poverty, and itappears that child care subsidies can greatly increase the chances that these motherswill work enough hours to improve their financial viability.

The relation between work and TANF receipt is negative and substantial, asexpected. The probability of an individual working 15 or more hours a week whenthey are not receiving TANF is 0.65. This figure drops to 0.40 when one receivesTANF. A similar drop is seen when we apply the full-time indicator. Here the prob-ability of working full time is only 0.48 when one is not receiving a subsidy, andthis figure drops all the way down to 0.25 when one is receiving TANF. For singlemothers with children under 5 the chances of working full time drop to 0.20 ifthey receive TANF. These figures, for both TANF and non-TANF mothers, high-light a complex and growing problem that our society is being forced to addressas the rate of single parent households increases. The probability that low-incomesingle mothers, under 35, will be in the workforce for 35 or more hours per weekis less than 0.50. This has serious implications for children growing up in thesesingle-parent families.

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Table 3. Predicted Probabilities (Calculated Holding All OtherVariables at Their Means)

Full Sample Single Mothers

Work Full time Work Full time

SubsidyNo 0.63 0.46 0.54 0.38Yes 0.79 0.60 0.75 0.53

Age16 0.56 0.38 0.51 0.3025 0.60 0.43 0.57 0.3835 0.65 0.49 0.64 0.4850 0.72 0.58 0.75 0.67

GenderNo (m) 0.85 0.76Yes (f ) 0.45 0.27

SingleNo (married/cohab) 0.61 0.44Yes 0.76 0.62

TANFNo 0.65 0.48 0.61 0.42Yes 0.40 0.25 0.34 0.20

Of the state-level characteristics included, only two, sophistication and workrequirements, reached statistical significance in any context. It is hypothesized thatstates with the most generous plans and the widest coverage will be positivelyrelated to employment. The variable only reaches statistical significance when con-sidering all parents of children under 5, in a context of working full time. Eventhen, the result has only minimal substantive significance. More troubling, the signis not in the predicted direction. The predicted probability of full-time employ-ment drops from 0.52 for a respondent in the least sophisticated states to 0.46 forthose in the most sophisticated states. It is not clear why the relation runs in thewrong direction. There could be a ceiling effect at work whereby states with thehighest sophistication have already reduced their rolls down to the hardest toemploy individuals.

The presence of work requirements meets statistical significance in the modelpredicting work for 15 or more hours per week within the full sample of parentswith children under 5. Again, the impact on the probability of working is a modest6-point difference. The predicted probability of working at least 15 hours in a statewithout work requirements is 0.60. When the parent lives in a state with assistanceconditioned on work requirements, the probability rises to 0.66. The correlationbetween work requirements and child care subsidies is very low, but it seems plau-sible that work requirements may serve as a proxy for a state’s commitment tomoving their low-income population into the workforce.

OLS Estimates

Taking advantage of the higher level of data available, the model is reestimatedusing the number of hours the respondent reported working per week in the pastyear as the dependent variable. The OLS estimates are presented in Table 4. Thepattern of statistical significance varies little from the logit models, but we do seea slight weakening of the significance of work requirements and sophistication.These two effects were already substantively weaker than child care subsidy, TANF,and the demographic variables. The R2 for the model is a modest 0.18, but thevariables of interest easily reach statistical significance, are correctly signed, andappear substantively important.

In the full model the child care subsidy variable has a highly significant coeffi-cient of 8.88, with a 95% confidence interval that ranges from 6.56 to 11.20 hoursper week holding all other variables constant. The same coefficient, when consid-ering the sample of single mothers, is 9.39 with a confidence interval ranging from6.69 to 12.1 hours per week. The coefficient is highly significant, with a standarderror of 1.38. Across all variations of the dependent variable, the child care subsidyvariable easily reaches statistical significance, and always moves in the hypothesizeddirection. Substantively, the values are encouraging, and they certainly point to acritical role for child care in helping low-income parents transition to work.4

Several attempts were made to improve upon the dichotomous subsidy measure,without success. It seemed that an interaction between subsidy receipt and statespending per child should be a strong indicator, with the underlying logic beingthat high-quality care would be correlated with more expensive care. By extension,we could hypothesize that mothers would be more willing to leave children for

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employment if they felt positively about the quality of care their child received whenthey were working. It also seemed plausible that parents would be more likely towork if they were required to spend less of their own wages on child care. To testthis hypothesis, subsidy receipt was interacted with the recipient’s state-requiredcopay level, set as a percentage of income for families 150% of the poverty line.Interestingly, neither of these interactions yielded significant results in relation towork levels. One possible explanation is that low-income parents cannot be as selec-tive as middle-income families when it comes to the quality of care and the deci-sion to work. If this is the case we would expect to see equal rates of employmentand effort between low-subsidy and high-subsidy holders, but we may find varia-tion in their level of satisfaction with the care their children receive. We may alsobe able to explain the lack of relation by adjusting for differences in the cost ofliving within states. It may be that high-subsidy states are those in which child carecosts are higher, which would reduce the value of their subsidies relative to otherstates. While this is plausible, it seems unlikely that this adjustment could fullyaccount for the wide variation in per child spending that we see across the states.In all models of employment, the only child care indicator that reached statisticalsignificance is the dummy of subsidy receipt. Only results from those models havebeen shown here.

The remaining individual-level variables all move in the expected direction andreach statistical significance. The substantive value of age is weaker than expectedwith a partial slope of 0.16 and a standard error of 0.05. Within the single motherssample the impact of age doubles to 0.36, or an average of one-third of an hourper additional year of age. This natural tendency to work less early in life, coupledwith the importance of earnings when raising a child, highlights the vulnerabilityof children born to young single mothers. In the full sample, the variable with thegreatest impact on the number of hours worked per week is gender. Holding all

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Table 4. OLS Coefficients

Full Sample Single Mothers

OLS-Hours Worked OLS-Hours Worked

Age 0.16* 0.36**(0.05) (0.08)

Gender −21.02** —(1.18)

Single 9.19** —(0.79)

TANF −12.02** −15.54**(2.08) (2.39)

Child care subsidy 8.88** 9.39**(1.18) (1.38)

Per capita income −0.000 −0.00(0.00) (0.00)

Sophistication 0.20 0.12(0.24) (0.36)

Work requirement 1.56* 0.40(0.82) (1.24)

Constant 35.18 21.05(3.78) (5.07)

Observations: 3,127 1,372

OLS/Logit coefficients, standard errors in parentheses; **p < 0.001; *p < 0.05.

other variables constant, women work on average 21 fewer hours per week thanmen do. Being single leads a person to work, on average, 9 hours more per weekthan the parent who is either married or cohabiting. This variable’s coefficient isalso highly significant, and has a confidence interval that ranges from 7.63 to 10.75hours per week. For many families this simply represents a value-based decision tokeep mothers at home while their children are young. For other families it is afinancially sound strategy depending on the number of children in the family andthe costs of available child care. Given the apparent reduction in hours workedassociated with being female and in a two-parent household, women who transi-tion to single-parent status will likely be required to undergo a dramatic shift intheir employment patterns if they are to achieve economic security. Unfortunately,having spent less time than their male counterparts in the workforce, they will lackthe economic advantages that experience and seniority bring.

The final individual-level variable in the model indicates whether or not therespondents, or their children, received TANF within the past year. Second togender, this variable has the largest substantive effect on the number of hoursworked per week. Of course, the causal arrow could easily move in the other direc-tion between these two variables. It is possible that to some extent receiving TANFactually reduces a person’s desire to work, but it is also probable that people firstwork fewer hours, and by way of low earnings then seek TANF. The partial slopefor the full sample of parents indicates that persons receiving TANF within the pastyear will on average work 12 hours less per week than those who are not receiv-ing assistance. The indicator has high statistical significance with a standard errorof 2.08, but a fairly wide confidence interval, which ranges from −16.11 to −7.92hours per week. The relation between TANF receipt and work effort is evenstronger in the sample of single mothers. Here the partial slope coefficient is −15.54 with a standard error of 2.39. The 95% confidence interval ranges from −10.85 to −20.24 hours per week.

The aggregate variables used to control for state variation are in all but one casenonsignificant.5 Per capita income and sophistication do not approach statisticalsignificance in either model, and sophistication moves in the wrong direction.Whether or not the state imposes work requirements does reach statistical signifi-cance in the full-sample specification, but the variable does not appear substan-tively important. The partial coefficient is 1.56 with a standard error of 0.83. Its95% confidence interval includes 0, and ranges from −0.057 to 3.19. It appearsthat the impact of work requirements on parents in this sample is minimal at best.Several other state policy measures were included in earlier models and in all butone case the results were nonsignificant.

Finally, a t-test is calculated to evaluate the significance of the difference betweenthe average number of hours subsidy recipients work and the average number ofhours worked by parents who sought a subsidy but did not receive one. The targetgroup consisted of the 448 parents in the sample that reported receiving help fromthe government with child care. These parents worked an average of 36 hours perweek, which exceeds the full-time threshold generally considered 35 hours perweek. The second group consists of the 99 subjects interviewed who reportedhaving requested help with child care but were either discouraged and gave up atsome point during the application process, or were placed on a waiting list until

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child care assistance became available.6 The average number of hours worked inthis group is 31, 5 hours less per week than those who receive subsidies, and lessthan full time. This test controls for a possible alternative explanation to increasedwork effort being attributable to child care assistance. It could be argued that thepeople who are motivated enough to seek this type of assistance are also more moti-vated to work in general than people who do not seek child care assistance. There-fore, we would expect people who are motivated enough to seek assistance to workmore regardless of whether or not they receive help, while those who do not seekhelp work less because they lack motivation more generally. We can test the influ-ence of motivation on work effort by comparing the average number hours workedby parents who received assistance and those who were similarly motivated butwere not granted a subsidy. The difference between the means is significant at the0.05 level, two tailed, with a t-score of 2.23. This suggests that among parents whoshare similar levels of motivation to work, those who receive assistance with childcare will tend to work more hours than those who do not. Parents in this samplewho received a child care subsidy, on average, worked full time while similarly moti-vated parents did not.

Discussion

The findings presented here suggest a need for further study on the impact ofchild care subsidies on employment patterns. This particular sample was restrictedto parents, yet substantial portions of the country’s poor children are cared for bynonparental family members (Rodgers, 2006, p. 151). The economic impact ofchild care subsidies could reach far beyond parent work effort and productivity.Future analyses should be opened up to include these nonparental caretakers.

The sample could also be extended in the future to parents with children over5 years of age. The challenge with this sample is that parents of school-age chil-dren generally face a lower financial child care burden than parents of childrenwho are not yet in the public education system. It is probably easier for parents tomake arrangements with family, friends, and older siblings for after-school carethan it is to find care for a full day. Still, summer breaks from school potentiallyimpact the employment patterns of many single parents, and subsidies could havean effect on the number of months out of the year parents of school-age childrenare employed.

We might also expect to see an increase in the impact of subsidies as the numberof young children in the home increases. While many child care providers offer areduction in rate for additional children in a family that are enrolled, for manyparents, and particularly single mothers, the cost of care for multiple children caneasily outweigh the benefit of working when subsidies are not available. Futureanalyses should control for the impact subsidies have on the work patterns ofparents with multiple children in care.

Finally, the NSAF dataset also provides rich information about the behavioral,emotional, and academic standing of the children and parents interviewed. Thisinformation could be modeled with subsidy receipt in an effort to estimate thenoneconomic benefits of providing child care subsidies. Additionally, the NSAF col-lected information regarding the setting in which children receive care, and to

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some extent the quality of care received. It would be possible to look at variationin the size of subsidies and how that relates to the type of care open to parents. Wecould then consider how the available care influences other family outcomes.Danziger, Ananat, and Browning (2004) investigate the influence of child caresubsidy on parent cost, parenting stress, and problems with care in a sample ofwelfare recipients in Michigan. They find substantial cost impacts but little impactin noneconomic domains. By expanding the scope of the study to include all low-income parents who are eligible for a subsidy, regardless of participation in otherprograms, the picture could change. For example, some modest support for thebroader benefit of child care has been seen in connection with Wisconsin’s NewHope reform effort (Bos et al., 2005; Mead, 2004). The evidence indicates thatunder certain conditions there are positive family outcomes associated withworking, and that children of parents who are working experience more rapid cog-nitive gains, lower levels of stress, and do better in school. Strengthening ourknowledge about the noneconomic benefits of moving low-income parents intowork, and their children into child care, is important to our overall understand-ing of the PRWORA’s impact.

Access to affordable child care is a critical prerequisite to work for many parents.As states transitioned their welfare systems to supported work programs followingPRWORA, it was clear that states would have to increase the scope and coverageof critical support services like child care subsidies. Along with the consolidation offour major child care funding sources into the Childcare and Development Fund,the amount of money available to spend on child care services increased five-foldbetween 1996 and 2003 (CCDF, 2003, p. 4). The multivariate analysis presentedhere lends support to the needed expansion of this critical support service. Parents,and particularly single mothers, are far more likely to work, and to work morehours per week when states help offset the costs of child care. Empirical evidenceindicates that the chances for escaping poverty are greatest when parents work atleast 35 hours per week (Rodgers, 2006, chap. 8). Getting these mothers into theworkforce full time is essential to raising their families out of poverty. The currentinvestigation suggests that receiving a child care subsidy moves single motherscloser to this goal.

Notes

1 Some earlier studies suggest that there may be a threshold effect whereby the number of hoursworked will not vary much in relation to subsidies but the decision to work or not will, so that receiv-ing a subsidy has an all or nothing effect on employment. Given the broad range of eligible parentsand provider types, I have chosen to reevaluate the impact on hourly effort. I use the dummy vari-ables initially to ease comparison between studies.

2 The question reads: In the past 12 months, did you receive government assistance in paying for childcare?

3 Blau and Tekin (2001) used additional control variables, but race was the only statistically significantone that was excluded from this model. I tried multiple variations on the race variable and all werenonsignificant. It is possible that as eligibility has opened up in the years since their analysis that therelation between race and subsidy receipt has weakened.

4 Blau and Tekin (2001) performed a two-stage least squares procedure to test for the possibility thatthe dependent variable influences welfare and child care subsidy receipt, and found no evidence thatindividuals who work more are more likely to seek out assistance (p. 29). Child care subsidy eligibil-

710 April Crawford

ity has been opened up considerably since their 1997 investigation, so the chances that work actuallyleads to subsidy is probably even smaller than it was in the past.

5 To see if there were important unmeasured state differences that had not been included in the model,I reran the model by state using STATA’s cluster function. The standard errors were almost identi-cal to the previous specification, suggesting that there are not unique state-contextual effects thathave gone unmeasured.

6 NSAF asks respondents whose income qualifies them for services that they are not receiving why theydo not receive the service. Respondent’s answers are coded as follows: did not want help, discour-aged, waiting list, and assistance not available. The group tested here were those respondents whosaid they had been discouraged or gave up, as well as those put on a waiting list.

About the Author

April Crawford is a graduate student in the Department of Political Science at the Univer-sity of Houston.

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