INSURANCE COVERAGE AT A POINT IN TIME
MEDICAID ELIGIBILITY AND PARTICIPATION
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1 PERCENTAGE DISTRIBUTION OF PARENT'S INSURANCE COVERAGE BY CHILD'S INSURANCE COVERAGE
2 DISTRIBUTION OF CHILDREN UNDER 19 BY DEMOGRAPHIC CHARACTERISTICS: SEPTEMBER 1994
3 DISTRIBUTION OF CHILDREN'S INSURANCE COVERAGE BY DEMOGRAPHIC CHARACTERISTICS: SEPTEMBER 1994
4 DISTRIBUTION OF CHILDREN'S DEMOGRAPHIC CHARACTERISTICS BY INSURANCE COVERAGE: SEPTEMBER 1994
5 DISTRIBUTION OF CHILDREN UNDER 19 BY SOCIOECONOMIC CHARACTERISTICS: SEPTEMBER 1994
6 DISTRIBUTION OF CHILDREN'S INSURANCE COVERAGE BY SOCIOECONOMIC CHARACTERISTICS: SEPTEMBER 1994
7 DISTRIBUTION OF CHILDREN'S SOCIOECONOMIC CHARACTERISTICS BY INSURANCE COVERAGE: SEPTEMBER 1994
10 MEDICAID PARTICIPATION RATES BY DEMOGRAPHIC CHARACTERISTICS: SEPTEMBER 1994
11 MEDICAID PARTICIPATION RATES BY SOCIOECONOMIC CHARACTERISTICS: SEPTEMBER 1994
12 CHILDREN UNINSURED IN SEPTEMBER 1994 AND EVER IN FY94, BY DEMOGRAPHIC CHARACTERISTICS
13 CHILDREN UNINSURED IN SEPTEMBER 1994 AND EVER IN FY94, BY SOCIOECONOMIC CHARACTERISTICS
24 ANNUAL FREQUENCY OF UNINSURANCE BY SINGLE YEAR OF AGE
25 ANNUAL FREQUENCY OF UNINSURANCE WITH MEDICAID ELIGIBILITY BY SINGLE YEAR OF AGE
26 ANNUAL FREQUENCY OF UNINSURANCE WITHOUT MEDICAID ELIGIBILITY BY SINGLE YEAR OF AGE
27 ANNUAL FREQUENCY OF MEDICAID COVERAGE BY SINGLE YEAR OF AGE
Using data from the Survey of Income and Program Participation (SIPP), this report examines the characteristics of children under the age of 19 by their health insurance status. The report presents findings on broad types of health insurance coverage while focusing in particular on Medicaid eligibility and participation and on children who lack coverage altogether--the uninsured. We include the following demographic characteristics: age, race and Hispanic origin, family composition, metropolitan versus nonmetropolitan residence, and region. We also include three socioeconomic characteristics: family income relative to the federal poverty level, parents' employment, and parents' education.
Section A presents findings with respect to health insurance coverage at a point in time. Section B examines the characteristics of children by Medicaid eligibility and participation. Section C looks at the lack of insurance over time. Section D presents estimates of the annual incidence and duration of uninsurance, Medicaid eligibility, and Medicaid participation by single year of age, utilizing an approach that yields the cleanest possible age differentials. Finally, Section E summarizes our major findings and conclusions.
In this section we examine the characteristics of children by their insurance coverage at a point in time, utilizing a measure of coverage that differentiates among employer-sponsored insurance, Medicaid, other public or privately purchased insurance, and the lack of coverage. We begin by comparing the insurance coverage of children with what was reported for their parents. We then examine how insurance coverage varies by children's demographic and socioeconomic characteristics.
1. Parent's Insurance Coverage
Table 1 presents a distribution of parent's insurance coverage by the child's insurance coverage for each of three points in time: October 1992, September 1993, and September 1994 (representing the beginning, middle, and end of the two-year period FY93 through FY94). "Parent" here refers to the individual identified in the SIPP file as the child's parent. Thus the parental coverage reported in Table 1 refers to only one parent. There are characteristics reported in later tables for which we take account of both parents, if present, where the second parent is identified as the spouse of the first parent. For health insurance coverage, however, we felt that we could not easily combine the four-category reports of two parents to produce a summary measure of both parents' coverage.
Altogether 18 to nearly 21 percent of uninsured children have a parent who reports some insurance coverage, and 7 to 8 percent of uninsured children have a parent who reports being covered by Medicaid. Another 4 percent of children appear to have no parent in the household, so their parents' insurance coverage must be regarded as unknown--and perhaps irrelevant.
We suspect that much if not most of the parent-child discrepancy in reported Medicaid coverage is due to measurement error--either in the misreporting of the child as not covered by Medicaid or the misreporting of the parent as covered. There are very few circumstances under which a parent would qualify for Medicaid while a child under 19 would not. If a parent is eligible for Medicaid by virtue of receiving Supplemental Security Income (SSI), then the children are not automatically eligible. It is likely, however, that many children who are in this situation would be eligible for Medicaid on the basis of their family income, given that the one parent receiving SSI and his or her SSI payments would not be counted in the Medicaid eligibility determination. Another situation that could result in a parent being covered by Medicaid but not the child is where the child qualifies for consideration as an adult. In this case the child could be ineligible for Medicaid despite having a
TABLE 1 (PDF document.)
parent and one or more siblings who qualify. We doubt that such cases as these could be numerous enough to account for the estimated 7 to 8 percent of uninsured children who have a parent covered by Medicaid, however.
If most of the reported instances of uninsured children with Medicaid-covered parents are indeed erroneous, we suspect that the reported lack of coverage of the child is where the problem lies. The Medicaid enrollment of children is underreported in the SIPP by 13 to 15 percent (see Technical Appendix D). Situations where Medicaid coverage is reported for a parent but not reported for all children may account for some of the underreporting among children. Certainly this possibility merits further investigation and, if borne out, could provide a basis for making a partial correction to the reported Medicaid coverage of children in the SIPP.
The 10 to 12 percent of uninsured children whose parents are reportedly covered by employer-sponsored insurance raise important questions as well. Could some of this, too, be due to error in the reporting of either the parent's coverage or the child's lack of coverage? If the parent is truly covered and the child is not, did the parent have no access to dependent coverage, or did the parent find it too expensive or otherwise choose not to obtain such coverage? These are not questions that SIPP data can answer, but they are questions that must be posed to other datasets and new data collection efforts if we are to understand the prevalence of uninsurance among children.
The 10 to 13 percent of Medicaid-covered children with uninsured parents is consistent with the expansion of Medicaid coverage aimed exclusively at children. That this percentage grows over time is also consistent with the newness of these expansions--participation in new programs tends to start slowly--and the fact that they have continued to make more and more children eligible over time.(1)
Finally, the consistent 6 percent of children with no parent in the household among children who are covered by Medicaid is noteworthy because it is three times as high as the rate at which all children under 19 appear to have no parents in the household. The rate among uninsured children is 4 percent or midway between the all-child rate and the Medicaid rate. This differential frequency of no-parent households among these different groups of children may reflect a causal impact of parental absence on both Medicaid participation and uninsurance. It is plausible, certainly, that the absence of parents from a child's household affects the likelihood that the child will qualify for Medicaid and, also, the likelihood that no private or other public insurance coverage will be provided.(2) At the same time, however, the absence of a parent from the survey household may reduce the likelihood that all sources of insurance coverage for a child are reported. In other words, the apparent relationship between a child's uninsurance and the parents' absence from the child's household may be due in part to the child's insurance coverage being underreported when neither parent is present in the household.(3) This might seem to be contradicted by the Medicaid finding, where children without parents are more likely than other children to be covered by Medicaid, but the mechanisms may be different.
2. Demographic Characteristics
In this section we examine differentials in children's health insurance coverage by five demographic characteristics: age, race and ethnicity (specifically, Hispanic origin), family composition, metropolitan residence, and region. The findings describe children in September 1994. Before presenting our findings we describe our measurement of the five characteristics and show how the population of children is distributed with respect to each of these characteristics.
a. Distribution of the Population
Table 2 presents the distribution of all children by the demographic characteristics that will be used to examine differentials in insurance coverage.
We divide children 0 through 18 into six age groups, using breaks that are relevant to Medicaid eligibility and, therefore, insurance coverage as well: infant, 1 to 5, 6 to 8, 9 to 10, 11 to 14, and 15 to 18. We separate the 9 to 10 group because these children phased into coverage under the poverty-related criteria over the two fiscal years that we examine. Specifically, none of the children aged 9 to 10 at the end of FY92 qualified for poverty-related coverage under the mandatory federal guidelines whereas all of these children met the age limit for poverty-related coverage by the end of FY94. Children 1 to 5 and 11 to 15 each account for 27 percent of the children under 19 represented by the SIPP panel in September 1994. Infants account for 4 percent while the other age groups include between 10 and 16 percent of all children.
We divide children into four categories on the basis of race and ethnicity. White non-Hispanic children account for 68 percent of the total. Black non-Hispanic children are 15 percent of the total while Hispanic children represent about 13 percent, and children of other racial groups--Asian and Pacific Islander and American Indian, Eskimo, and Aleut account for the remaining 4 to 5 percent.
TABLE 2 (PDF document.)
We define family composition in terms of which parents are present in the household. Children living with both parents account for 72 percent of all children while children with only their mothers present represent 24 percent. The remaining four percent are divided evenly between children living with their fathers and children with no parent present. The latter will include older teens living on their own or with friends as well as younger children being raised by relatives or others who are not their natural or adoptive parents.
The SIPP files do not identify rural residents explicitly, so we have used metropolitan area residence to differentiate between a group including primarily urban children and a group including primarily rural children. The metropolitan category includes 74 percent of children while the non-metropolitan category includes 25 percent. The SIPP files assign 1.2 percent of children to a "not applicable" category, which is a Census Bureau designation that is not defined.
We used the census region classification to break down children by region of the country. The nine census regions and the states they include are:
New England: Maine, New Hampshire, Vermont, Massachusetts, Rhode Island, and Connecticut
Middle Atlantic: New York, New Jersey, and Pennsylvania
East North Central: Ohio, Indiana, Illinois, Michigan, and Wisconsin
West North Central: Minnesota, Iowa, Missouri, North Dakota, South Dakota, Nebraska, and Kansas
South Atlantic: Delaware, Maryland, District of Columbia, Virginia, West Virginia, North Carolina, South Carolina, Georgia, and Florida
East South Central: Kentucky, Tennessee, Alabama, and Mississippi
West South Central: Arkansas, Louisiana, Oklahoma, and Texas
Mountain: Montana, Idaho, Wyoming, Colorado, New Mexico, Arizona, Utah, and Nevada
Pacific: Washington, Oregon, California, Alaska, and Hawaii
While the SIPP file does not individually identify nine of the 50 states, leaving us to randomly assign children in each of three groups to individual states, only one of these nine states is not grouped with others from the same region.
Four of the nine regions include between 14 and 18 percent of the child population, with the East North Central region being the largest. Two of the regions--New England and the Mountain region--include less than 5 percent of children under 19. The remaining three account for 7 to 11 percent of all children.
b. Distribution of Insurance Coverage
Table 3 presents for each of the five demographic characteristics the distribution of insurance coverage by subgroup.
The inverse relationship between age and Medicaid coverage and the direct relationship between age and uninsurance are striking, but it is important to keep in mind the generally antecedent role of employer-sponsored and other insurance. The older the child, the more likely that the child is covered by either employer-sponsored or other insurance (generally privately purchased or public insurance other than Medicaid). The 71 percent of children 16 to 18 who are covered by either of these sources contrasts with 64 percent of infants and 65 percent of children 1 to 5. Among those children who might otherwise be uninsured, Medicaid is much more likely to be utilized by younger children (who are more likely to be eligible) than older children. At the extremes, Medicaid covers nearly three-quarters of the infants who lack other coverage but less than half of the 16- to 18-year- olds who, similarly, have no other coverage. The percentage of all children covered by Medicaid in each age group varies from 12 percent among children 16 to 18 to a maximum of 26 percent among infants, with the proportion rising monotonically as age decreases. The frequency of uninsurance, conversely, rises from 9 percent among infants to 16 percent among children 16 to 18.
TABLE 3 (PDF document.)
There are very strong differentials by race and ethnicity (Hispanic origin). Non-Hispanic whites have the highest rate of coverage by employer-sponsored plans at 74 percent and the lowest rates of Medicaid coverage (about 11 percent) and uninsurance (10 percent). Non-Hispanic blacks have the highest rate of Medicaid coverage at 44 percent or four times the rate among whites. Blacks also have the second lowest coverage by employer-sponsored plans at 42 percent, but the rate of uninsurance among blacks is only 2.5 percentage points above that of whites. Hispanics report the highest rate of uninsurance by far at 26 percent, with the lowest rate of coverage by employer-sponsored plans, but 34 percent report Medicaid coverage--a rate that is three times that of non-Hispanic whites. In all coverage groups but other insurance the other races category falls between non-Hispanic whites and Hispanics--generally closer to whites. While coverage by other than employer-sponsored plans or Medicaid is quite low at 4 percent nationally, non-Hispanic whites and other races report much higher coverage--between 5 and 6 percent--than do Hispanics and non-Hispanic blacks.
Differentials by family composition are as great as those by race and ethnicity. Children with both parents present have the highest rates of employer-sponsored and other coverage at 75 percent and 5.0 percent, respectively, and the lowest rates of Medicaid coverage and uninsurance: 8.5 percent and about 12 percent. At the opposite end, children with no parent present have the lowest rate of employer coverage at 27 percent--barely one-third the rate for children with both parents present--and the highest rate of uninsurance at 23 percent or double that of children with both parents present. Children with no parent present share the highest rate of Medicaid coverage with children in mother- only families, at 47 to 48 percent. Children in mother-only families have a much lower rate of uninsurance than children in father-only or no-parent families. Compared to children with no parent present, children in mother-only families have a higher rate of employer-sponsored coverage. Compared to children in father-only families, children in mother-only families are covered by Medicaid at a rate that more than compensates for their lower employer-sponsored coverage.
Differences by metropolitan residence are marginal at best. Children in non-metropolitan areas have a higher rate of uninsurance by two percentage points. Their lower rate of employer-sponsored coverage is partially offset by marginally higher coverage rates for Medicaid and other insurance. Children who could not be classified as metropolitan or not are too few in number for their coverage rates to provide meaningful information, but their rates of employer-sponsored coverage and uninsurance are strikingly similar to those of non-metropolitan children.
Finally, regional differences in rates of uninsurance are quite substantial. New England has the lowest rate at 7.1 percent while the West South Central region has the highest rate at 22 percent or three times the rate observed in New England. In general, rates of uninsurance are highest in the South and the West. Much of the latter, we suspect, can be attributed to the high proportion Hispanic in the Western states. The role of Medicaid in these regional differences in rates of uninsurance is ambiguous--no doubt because higher than average rates of Medicaid participation can reflect either or both the prevalence of poverty and broader eligibility criteria. Employer-sponsored coverage varies over a range of 20 percentage points, with New England having the highest rate of coverage and the Pacific states having the lowest, followed closely by the East and West South Central states.
c. Distribution of Demographic Characteristics within Coverage Groups
Table 4 is a recasting of Table 3 with column percentages instead of row percentages. For each demographic characteristic it shows the percentage distribution of demographic subgroups within each class of insurance coverage. Distributions of demographic subgroups within coverage groups are of interest insofar as they indicate that the members of a particular coverage group--such as the uninsured--have a markedly different demographic composition or profile than other coverage groups. Such distributions are of value in determining how much of a specific deficiency in coverage could be addressed by a targeted strategy focusing on certain subgroups. In general, sizable differentials in coverage by demographic or other characteristics imply sizable differences in the profiles of coverage groups--and vice versa. When coverage differentials are confined to differences between small subgroups and the balance of the population, however, the profiles of different coverage groups may appear very similar. The absence of large differences in the profiles of the various coverage groups does not imply that the differentials are unimportant, necessarily, but it does indicate that a narrowly targeted policy will not have much impact on overall coverage.
Age distributions differ little across the coverage groups except for Medicaid, where children under 6 account for 42 percent of children reporting Medicaid coverage but only 25 to 29 percent of each of the other three insurance classes. Clearly the age profile of Medicaid enrollees is different from that of children with other sources of coverage--or no coverage at all--but these other coverage groups differ little from each other in their age composition. Racial composition varies substantially across the coverage groups, however. Hispanic children represent 26 percent of the uninsured and 23 percent of Medicaid enrollees but only 8 percent of children covered by employer-sponsored plans. Non-Hispanic blacks account for 35 percent of Medicaid enrollees but less than 15 percent of the uninsured and 10 percent of children with employer-sponsored plans. Non-Hispanic white children dominate every group except Medicaid enrollees. They account for more than half of the uninsured and more than three-quarters of those with employer-sponsored coverage but only 38 percent of the Medicaid population.
TABLE 4 (PDF document.)
The impact of group size is evident for family composition as well. While children in two-parent families have the lowest incidence of uninsurance, they nevertheless account for 65 percent of all uninsured children. Children from mother-only families dominate Medicaid due to the eligibility rules for families. They account for 60 percent of Medicaid enrollees compared to only 13 percent of those with employer-sponsored coverage. The father-only and no-parent groups do not account for more than 6 percent of any coverage group.
The only group in which children from non-metropolitan areas account for noticeably more than their share of all children is other insurance, where they represent 34 percent of the total versus 24 to 28 percent for other coverage groups. Finally, while individual regions account for varying shares of the different coverage groups, the variation across coverage groups is generally modest. Perhaps the most notable exception occurs in the West South Central region, which accounts for less than 9 percent of Medicaid enrollees but more than 18 percent of the uninsured.
3. Socioeconomic Characteristics
Differentials in children's insurance coverage by key socioeconomic characteristics are as sizable as those we have reported for demographic characteristics. Here we examine differentials by family poverty level, parents' employment, and parents' education as observed in September 1994. We begin by describing our measurement of the three characteristics and showing how children under 19 are distributed by each of the three. Following this we present distributions of insurance coverage by levels of each of these characteristics and then report the percentage distribution of children by poverty level, parents' employment, and parents' education within each category of insurance coverage.
a. Distribution of the Population
To calculate the family poverty level for each child, we used the September 1994 family income and family size, as constructed by the Census Bureau, and compared the family income to 1/12 the annual poverty threshold for a family of the size reported in that month.(4) By this measure, 9 percent of children under 19 were in families below 50 percent of the poverty line, and nearly 12 percent were in families between 50 and 100 percent of poverty, yielding a total poverty rate close to 21 percent (see Table 5).(5) A slightly larger percentage of children were in families between 100 and 200 percent of poverty while a slightly smaller fraction were in families between 200 percent and 300 percent of poverty. Just over one-third or 35 percent of children were in families above 300 percent of poverty.
We measured parents' employment by first determining whether each parent present in the child's household in September 1994 was employed full time, part time, or not at all in that month, and then we classified children as having one or both parents employed full time, neither parent employed full time but one or both parents employed part time, and neither parent working. Children with no parent identified as present in the household were classified separately. By this measure, 78 percent of children had one or both parents employed full time, 5 percent had parents employed only part time, 14 percent had no parents working, and 2 percent had no parents present in the household.
TABLE 5 (PDF document.)
To classify children by their parents' education, we assigned each parent present in the child's household in September 1994 to one of six educational levels, based on years of schooling completed and whether the parent obtained a college degree. We then classified each child by the higher of either parent's level of education when both were present or by the one parent's level of education when only one was present. Children with no parent present were assigned to a separate category. Only 2 percent of children had parents with six or fewer years of schooling, and 10 percent had parents with at most 7 to 11 years of schooling. The largest group of children, at 33 percent, had at least one parent with exactly 12 years of schooling (and the other, if present, with 12 years or less). The remaining 52.5 percent of children had parents who attended college but did not graduate (24 percent), completed a four-year degree (15 percent), or went on to graduate school (about 14 percent).
b. Distribution of Insurance Coverage
Table 6 provides estimates of differential insurance coverage by family poverty level, parents' employment, and parents' education.
As we would expect, both employer-sponsored coverage and Medicaid are strongly related to the child's poverty level, but in opposite directions. Only 11 percent of children in families below 50 percent of poverty and 21 percent of those between 50 and 100 percent of poverty have employer-sponsored coverage. More than half of those in families between 100 and 200 percent of poverty have such coverage, and this proportion rises to 90 percent among children in families above 300 percent of poverty. For Medicaid we find that 70 percent of children below 50 percent of poverty and 53 percent of those between 50 and 100 percent of poverty are reported to be covered. Medicaid coverage declines to 20 percent among those between 100 and 200 percent of poverty then plunges to 5 percent for those between 200 and 300 percent of poverty and drops further to below 2 percent among children at 300 percent of poverty or higher.
TABLE 6 (PDF document.)
Other insurance shows little relationship to poverty level. Children in families below poverty are somewhat less likely to have other insurance coverage, but there is no variation among children above poverty. Uninsurance is inversely related to poverty level but not linearly: children below 50 percent of poverty have a lower rate of uninsurance than children between 50 percent and 200 percent of poverty. Clearly, the Medicaid coverage that is available to children below 50 percent of poverty explains this group's comparatively low rate of uninsurance.
Parents' employment has a very strong relationship with the child's insurance status as well. Children with at least one parent working full time have the highest rate of employer-sponsored coverage at 77 percent, the lowest rate of Medicaid coverage at 7.3 percent, and the lowest rate of uninsurance at 11 percent. Employer-sponsored coverage plunges to 9 percent, and Medicaid coverage rises to 73 percent when there is no working parent in the household. Uninsurance rises as well but to a level between that of children with parents employed full time versus part time. Children with no parent present have rates of both employer-sponsored coverage and Medicaid coverage that are between those of children whose parents work part time and children whose parents do not work at all, but they have the highest rate of uninsurance, at 23 percent.
Parents' education is highly related to the incidence of every one of the four coverage groups. Employer-sponsored coverage increases from 11 percent to 86 percent from the lowest education level (six years or less) to the highest (some graduate work). Medicaid coverage declines from 46 percent to 3 percent between the lowest and highest levels of education although children with parents educated only 7 to 11 years have a higher Medicaid participation rate (53 percent) than those with less educated parents. Other insurance coverage grows from 0 percent to 7 percent between the lowest and highest levels of education while the rate of uninsurance declines from 43 percent to 4 percent. For each of these four coverage groups this is by far the broadest range of rates seen for any of the demographic or socioeconomic variables. Given the nonlinear relationship between family poverty level and a child's lack of insurance, we speculate that the strength of the relationship between parents' education and their children's lack of insurance may reflect more than just differential access to health insurance. Certainly this merits further investigation, including the examination of education differentials in a multivariate context, but if the relationship holds up there are clear policy implications for efforts to increase enrollment in Medicaid and the new state Children's Health Insurance Program (CHIP) initiatives.
c. Distribution of Socioeconomic Characteristics within Coverage Groups
Table 7, which is analogous to Table 4, gives the percentage distribution of children by poverty level, parents' employment, and parents' education within each class of insurance coverage.
Children in families between 100 and 200 percent of poverty account for 39 percent of the uninsured. The size of this portion of the uninsured is important because it includes the children addressed most explicitly by the CHIP legislation. Older children between 50 and 100 percent of poverty also constitute a prime target of CHIP. Across all ages this group accounts for 21 percent of the uninsured. Somewhat surprisingly, however, uninsured children in families above 300 percent of poverty, at 10 to 11 percent of all uninsured children, are nearly as numerous as uninsured children in families below 50 percent of poverty, who account for 12 percent of the total.
It is particularly clear from Table 7 how Medicaid complements employer-sponsored and other insurance coverage. Children below 100 percent of poverty account for only 5.3 percent of all children with employer-sponsored coverage and 13 percent of those with other coverage, but they account for 67 percent of the children covered by Medicaid. Children between 100 and 200 percent of poverty account for relatively similar shares of all children covered by employer-sponsored plans, Medicaid, or other insurance. Children above 200 percent of poverty represent 75 percent of the children covered by employer-sponsored plans and 58 percent of the children covered by other insurance but only 9 percent of the children covered by Medicaid.
TABLE 7 (PDF document.)
With respect to parents' employment what stands out is the 69 percent of uninsured children who have at least one parent employed full-time. At the same time, children with parents working part-time account for only 19 percent of the uninsured. Children with no working parents account for 18 percent of the uninsured but 54 percent of those enrolled in Medicaid.
Children whose more educated parent had exactly 12 years of schooling constitute one-third of all children under 19, but they represent 42 percent of children enrolled in Medicaid and 42 percent of the uninsured. Interestingly, the uninsured are distributed almost identically below and above this modal educational level whereas Medicaid enrollees come disproportionately from children whose parents had less than 12 years of education. This latter group contributes only 4 percent of children with employer-sponsored coverage and 1 percent of those with other coverage
The number of uninsured children who appear to be eligible for Medicaid, according to survey estimates, has raised concerns about the adequacy of Medicaid outreach. In Appendix B we presented findings from the 1992 SIPP panel that suggest that: (1) most uninsured children who become eligible for Medicaid remain eligible for very short periods of time, (2) more than two-thirds of all spells of Medicaid-eligible uninsurance are preceded or followed by periods of uninsurance without Medicaid eligibility, and (3) about one-third of spells of Medicaid-eligible uninsurance are preceded and followed by spells of uninsurance without Medicaid eligibility. The often transitory nature of Medicaid eligibility during spells of uninsurance may help to explain why many uninsured children who appear to be eligible for Medicaid do not enroll. In addition, we found that more than 40 percent of spells of Medicaid-eligible uninsurance were preceded or followed by Medicaid enrollment, and perhaps one fifth of such spells were preceded and followed by Medicaid enrollment. In other words, a significant share of Medicaid-eligible uninsured children appear to have initiated their spells of uninsurance by leaving Medicaid or to have ended their spells of Medicaid-eligible uninsurance by enrolling in Medicaid. For the first group, the key policy question is not why these children have not enrolled in Medicaid but, rather, why they left Medicaid. For the second group the key policy question is not why have these children not enrolled in Medicaid but why have they not enrolled sooner. Thus our analysis of the dynamics of uninsurance, Medicaid eligibility, and Medicaid participation suggests that inferences about the inadequacy of Medicaid outreach from the observation that three or four million uninsured children appear to be eligible for Medicaid at any point in time are overly simplistic and may be only partially supported by a closer examination of this population--particularly before and after the period of Medicaid eligibility. In the first subsection below we examine the demographic and socioeconomic characteristics of uninsured children who appear to be eligible for Medicaid, and in the second subsection we look at Medicaid participation rates among all eligible children by their demographic and socioeconomic characteristics.
1. Medicaid Eligibility among Uninsured Children
Table 8 looks at Medicaid eligibility among uninsured children by demographic characteristics, and Table 9 does so by socioeconomic characteristics. Each table presents the number uninsured and the percent of these who are Medicaid-eligible for each demographic or socioeconomic group.
TABLE 8 (PDF document.)
TABLE 9 (PDF document.)
Each table also indicates what share of the Medicaid-eligible uninsured and the Medicaid-ineligible uninsured each demographic or socioeconomic group represents.
Overall, 33 percent of the uninsured were estimated to be eligible for Medicaid in September 1994.(6) The sharpest differentials in this percentage are by age of child. The eligibility rate is 68 percent among infants, then drops to between 42 percent and 48 percent among children age 1 to 10, then drops further to 18 or 19 percent among older children. Children 1 to 5 represent 32 percent of Medicaid-eligible uninsured children while children 6 to 10 combine to represent 34.5 percent of this population. Infants represent 6 percent of the Medicaid-eligible uninsured. Children 11 to 15 represent less than 17 percent of the Medicaid-eligible uninsured but 35 percent--the largest share--of children who are uninsured but not Medicaid-eligible. Similarly, children 16 to 18 account for 24 percent of the Medicaid-ineligible uninsured but only 10 percent of the Medicaid-eligible uninsured.
Racial and ethnic differences in the proportion of uninsured children who are eligible for Medicaid reflect differences in family composition and income level. Non-Hispanic blacks have the highest eligibility rate at 45 percent despite the fact that blacks also have the highest rate of Medicaid coverage by far (Table 3). Hispanic children and those of races other than white or black have eligibility rates between 37 and 40 percent while non-Hispanic whites have a 27 percent eligibility rate. Because of their overall numbers, non-Hispanic whites account for the largest shares of both the Medicaid-eligible and Medicaid-ineligible uninsured at 44 percent and 58 percent, respectively. Hispanic children account for the next largest shares of both groups: 29 percent of the Medicaid-eligible and 25 percent of the ineligible uninsured.
Nearly 51 percent of uninsured children in mother-only families are eligible for Medicaid compared to 33 percent in father-only families, 27 percent in two-parent families and 17 percent among children living without their parents. Again, because of size, two-parent families account for the largest shares of both the Medicaid-eligible and ineligible uninsured at 52 percent and 71 percent respectively. Children in mother-only families represent 42 percent of the Medicaid-eligible uninsured and 20 percent of the Medicaid-ineligible. Children in father-only or no parent families account for 5 percent or less of either group.
Uninsured children living in metropolitan areas are somewhat more likely to be eligible for Medicaid than those living in non-metropolitan areas (35 versus 28 percent), and they account for 74 percent of the Medicaid-eligible uninsured and 69 percent of the Medicaid-ineligible uninsured.
Except for New England, the regions vary by only 11 percentage points in the proportion of their uninsured children who are Medicaid-eligible. Children in the West South Central region have the lowest eligibility rate at 27 percent while uninsured children in the two North Central regions have a 38 percent Medicaid eligibility rate, which is 10 percentage points below that of New England. Consistent with this fairly limited variation in Medicaid eligibility, each region's share of all Medicaid-eligible uninsured children is fairly comparable to its share of Medicaid-ineligible uninsured children.
Medicaid eligibility among uninsured children is highly related to family income level, as we would expect. We see in Table 9 that uninsured children below 50 percent of poverty have a 78 percent Medicaid eligibility rate. Family composition, family resources, and low eligibility thresholds in some states account for why this eligibility rate is not even higher. The Medicaid eligibility rate among children between 50 and 100 percent of poverty is 55 percent while the rate among children between 100 and 200 percent of poverty is about half that level. Children in each of the three lowest poverty classes account for similar shares of all the Medicaid-eligible uninsured: between 28 and 35 percent. Children in families between 100 and 200 percent of poverty account for the largest share of uninsured children who are not eligible for Medicaid--41 percent--while children in families between 200 and 300 percent of poverty represent 25 percent of the Medicaid-ineligible uninsured.
Parents' employment is inversely related to uninsured children's Medicaid eligibility. Uninsured children with at least one parent working full time have a 24 percent Medicaid eligibility rate compared to 44 percent for children with parents working only part time. Children with no working parent have a 64 percent Medicaid eligibility rate, and they account for 35 percent of all Medicaid-eligible uninsured children. Children with at least one parent working full time, who represent more than two-thirds of all uninsured children, account for 50 percent of the Medicaid-eligible uninsured and 78 percent of the ineligible uninsured. Children with no parent present have the lowest Medicaid eligibility rate at 17 percent and account for only 2 percent of the Medicaid-eligible uninsured and 5 percent of the ineligible uninsured.
Medicaid eligibility varies unevenly by parents' education. Children whose parents have at most 7 to 11 years of schooling have the highest Medicaid eligibility rate at 43 percent while children whose parents are college graduates have the lowest eligibility rate at 24 to 25 percent. Children whose parents have exactly 12 years of schooling have a Medicaid eligibility rate matching the national average but because of their numbers account for 43 percent of all Medicaid-eligible uninsured children and 41 percent of Medicaid-ineligible uninsured children. Shares of both populations decline as parents' education rises or falls from this median level.
Interpretation of these very high rates of Medicaid eligibility that we see among subgroups of the uninsured must be tempered by some of our findings from the examination of eligibility over time, which suggest that spells of Medicaid-eligible uninsurance frequently end with children enrolling or re-enrolling in Medicaid or by losing their eligibility, and that spells of Medicaid-eligible uninsurance typically last just a few months. Furthermore, it is important to recognize that Medicaid enrollment of children may be underreported by more than 2 million children at a point in time (see Technical Appendix D), which is about the number of uninsured children that we estimate to be eligible for Medicaid. Now, not all of the children whose Medicaid enrollment is unreported will have been reported as uninsured, and our simulations underestimate the number of children who are eligible for Medicaid. Nevertheless, it is undoubtedly correct to infer that the Medicaid eligibility rate among uninsured children is overstated in Tables 8 and 9. What implications this has for the observed differentials is not clear, however. It is plausible that Medicaid underreporting is greatest for subgroups with the highest percentages of Medicaid eligibles among their uninsured. If so, the differentials themselves may be overstated. On the other hand, if the underreporting of Medicaid coverage is due in large part to a stigma that respondents associate with participation, then respondents who belong to subgroups with low participation could feel this stigma most strongly, with the result that Medicaid participation is more likely to be unreported in subgroups with low participation than in subgroups with high participation.
2. Medicaid Participation Rates
Tables 10 and 11 present Medicaid participation rates in September 1994 by demographic and socioeconomic characteristics among all eligible children and among children with no other
TABLE 10 (PDF document.)
TABLE 11 (PDF document.)
insurance coverage. The first participation rate is defined as the number of Medicaid-enrolled children who were simulated to be eligible divided by the number of all children who were simulated to be eligible.(7) The second participation rate is defined as the number of Medicaid-enrolled children who were simulated to be eligible divided by the number of children who were simulated to be eligible and had no other source of insurance. That is, the participation rate is calculated as the number of Medicaid-enrolled children who are simulated to be eligible divided by the sum of these same Medicaid-enrolled children and Medicaid-eligible uninsured children.
Calculated over all eligible children, the estimated Medicaid participation rate is about 65 percent. When we exclude from the denominator those eligible children who are covered by insurance other than Medicaid, the estimated participation rate rises by 14 percentage points to 79 percent.(8) Neither rate includes an adjustment for the underreporting of Medicaid participation in the SIPP, and, as explained above, neither rate includes reported participants who were simulated to be ineligible. Even with these caveats, which imply that the rates may be understated, the participation rate among children who would otherwise be uninsured is quite high. While the participation rate among all eligible children is significantly lower, the fact that the difference between the two rates is attributable solely to children who were reported to have other coverage puts the overall participation rate in a new light. Certainly we should not assume that this other coverage was necessarily more comprehensive than Medicaid, and in many cases almost surely it was not. But for a significant number of children who were Medicaid-eligible but not participating the choice faced by their parents appears to have been between Medicaid and another source of coverage rather than between Medicaid and no insurance at all. Why parents in this situation may have elected not to enroll their children in Medicaid is a question whose answer may have implications for Medicaid and CHIP outreach as well as for strategies to minimize the potential crowd-out effects of CHIP.
Comparison of the two columns of Table 10 shows that in addition to raising the participation rate by 14 percentage points, on average, the effect of excluding children with other insurance from a Medicaid participation rate is to render the rates much more equal across demographic subgroups. This is most striking for race and ethnicity, where non-Hispanic whites lag behind Hispanics by 13 percentage points and behind non-Hispanic blacks by 25 percentage points when the participation rate is calculated among all Medicaid-eligible children. When children with other insurance are excluded from the participation rate, whites show effectively the same participation rate as Hispanics (75 percent versus 74 percent) and lag behind blacks by only 13 percentage points. Puzzling age differences are largely eliminated when the rate is calculated just among children who would otherwise be uninsured, and differentials by family composition are reduced as well. A small metropolitan/non-metropolitan difference is eliminated, and the surprisingly low participation rate in New England is increased by 24 percentage points with the alternative measure. While regional differences remain, only one region--the West South Central--is very far out of line with the rest with a participation rate of 69 percent or 19 percentage points below the highest participation rate. In fact, if these two regions are eliminated, the remaining seven regions fall within a 5 percentage point range.
The story is much the same for socioeconomic differentials (Table 11). When all eligible children are included in the calculation, the range of participation rates among poverty classes is 31 percentage points. This drops to 20 percentage points when children with other insurance are excluded, and the relationship between poverty and Medicaid participation becomes U-shaped, with the lowest participation rate found in the middle poverty group and the highest rates found at the ends. Parents' employment shows a similar equalizing of rates with the alternative measure of participation. There remains a monotonic, inverse relationship between the level of parents' employment and their children's participation rate in Medicaid, but the range of participation rates is reduced from 46 percentage points to 26 percentage points. Finally, differentials in children's participation rates by parents' education level are reduced as well, with a 20 point differential across the middle three levels being reduced to 4 percentage points. The children of college graduates (including those who did graduate work) remain much less likely to participate than the children of less educated parents, but instead of being 45 percentage points below the group with the highest participation rate they are less than 25 percentage points below this group. Nevertheless, we continue to see evidence suggesting that the education differential reflects more than just access to health insurance. The 70 percent participation rate for children whose parents have no more than an elementary school education contrasts with participation rates of 84 percent for children below 50 percent of poverty and 86 percent for children with no working parent. The comparatively low participation rate of children in this lowest parents' education category underscores the need for further research on the relationship between education and insurance coverage.
Why do the differentials in participation rates decline when we exclude children with other coverage, and what does this tell us about participation in Medicaid? Some compression of the differentials may have occurred simply because we reduced the variance of participation by removing a group of nonparticipants from the rates. The dominant reason for the reduction, however, is that in removing children with other sources of coverage we have removed one source of the original differentials. That is, differences in Medicaid participation rates exist in part because children in different demographic and socioeconomic groups are differentially likely to have other coverage. That we observe these reductions in differential participation rates also suggest that it is unlikely that most of this other coverage is misreported Medicaid coverage. If it were largely Medicaid, then participation rates would be boosted more uniformly--unless, of course, the misreporting itself occurred disproportionately among groups that were least likely to participate. This is not implausible--particularly if the perceived stigma attached to Medicaid contributed to the misreporting, as well it might. The Medicaid stigma might also contribute to parents' decisions to choose an alternative source of coverage over Medicaid if such coverage is available. Clearly, we have identified an area where further research would be beneficial, as there could be important policy implications in the reasons why parents may have and choose alternatives to Medicaid, as we noted previously.
From looking at differentials in health insurance coverage at a point in time we turn now to differentials in the lack of health insurance coverage over time. First we compare children who were uninsured at the end of a year with those who were ever uninsured during the year. Next we look at differentials in the duration of all new spells of uninsurance that began during FY93, followed by new spells of uninsurance coupled with Medicaid eligibility. We close by examining differentials in the duration of spells that were active at the end of FY93 and comparing these to differentials in the duration of new spells that began during the year.
1. Uninsured At a Point in Time Versus Ever in a Year
Table 12 compares demographic differentials in the number of children uninsured in September 1994 and the number ever uninsured during FY94, and Table 13 compares socioeconomic
differentials. In both tables we see that the percentages of children ever uninsured during the year are much higher than the percentages uninsured in September 1994, and this in itself is striking, but the differentials are not materially different except for race/ethnicity and region. Black non-Hispanic children are more likely than children in the "other" group of races to have been ever uninsured during the year but less likely to be uninsured in September 1994. With respect to region there is a very clear division between the Northeast and Midwest, on the one hand, and the South and West on the other in the percentage ever uninsured during FY94. In the Northeast and Midwest the proportions ever uninsured during the year vary from 12 to 17 percent while in the South and West these proportions range between 22 and 33 percent. We note as well that the very large differentials in the frequency of uninsurance by parents' education are even more impressive with respect to children ever uninsured during the year. The proportion of children ever uninsured in FY94 rises from 8 percent among children whose parents did some graduate work after college to 54 percent among children whose parents received less than 7 years of education
Consistent with the generally very limited change in differentials between children uninsured in September 1994 and children ever uninsured during FY94, shares of the two groups of uninsured show little difference. This holds true even where the differentials in the proportion ever uninsured during the year are clearly larger than the differentials in the proportion uninsured in September 1994. For example, while the difference between Hispanic children and non-Hispanic white children grows from 16 percentage points to 20 percentage points, the Hispanic share of children who were ever uninsured is actually 3 to 4 percentage points lower than the Hispanic share of children
TABLE 12 (PDF document.)
TABLE 13 (PDF document.)
uninsured in September 1994. Because the white population is so much larger than the Hispanic population, a smaller percentage point change can nevertheless add more white children than Hispanic children to the number of uninsured. We see a similar phenomenon with respect to region. Despite larger percentage point increases in the proportion uninsured in the South and West, the share of uninsured children living in these regions is no higher for children ever uninsured in FY94 than for children uninsured in September 1994.
2. Duration of New Spells of Uninsurance
We have demonstrated that there are very pronounced demographic and socioeconomic differentials in the likelihood that a child is uninsured at a point in time or over the course of a year. Do these differences extend to the features of children's spells of uninsurance? In particular, are there important differences in the duration of spells of uninsurance by demographic or socioeconomic characteristics? We examined differentials in the distribution of completed spells of uninsurance among spells that started in FY93. Our findings with respect to demographic characteristics are presented in Tables 14 and 15. Socioeconomic differentials are reported in Tables 16 and 17.
As we explained in Appendix B, the durations of completed spells of uninsurance as measured in the SIPP incorporate a significant type of measurement error that greatly overstates the reported frequency of spells lasting exact multiples of four months. Consequently, the distributions reported in the tables that follow should not be interpreted literally. In particular, the 54 percent of all new spells that are reported as having been completed in 1 to 4 months is probably overstated, as some of the many spells identified as being completed in exactly 4 months undoubtedly lasted 5, 6, or even 7 months. Similarly, the percentage of spells lasting 13 months or more is probably understated, given the likely overreporting of spell durations of exactly 12 months. Intermediate spells are
TABLE 14 (PDF document.)
TABLE 15 (PDF document.)
TABLE 16 (PDF document.)
TABLE 17 (PDF document.)
understated at the low end and overstated at the high end, with an uncertain effect on their overall frequency.(9)
Despite this measurement error in the reported durations of spells, differentials in the relative frequency of spells of different length may carry important information about differences in the experience of uninsurance among children across demographic and socioeconomic subgroups. We recognize, though, that measurement error in the reporting of durations may vary by these same characteristics and either weaken or distort the observed relationships.
The results reported in Table 14 suggest that, generally, demographic differentials in the reported duration of children's spells of uninsurance are weak at best. There are no meaningful differences by age. With respect to race and ethnicity we find that Hispanic children have the fewest spells of 1 to 8 months in length and the most spells of 13 months or longer, but the differences between Hispanic children and other children are modest. Turning to family composition, we find that children in mother-only and no parent families have the highest incidence of spells reported to have lasted 4 months or fewer while children in mother-only families also have the lowest incidence of spells exceeding 12 months. There are no differences by metropolitan residence but there are differences by region. Spells tend to be shortest in New England and longest in the East South Central and Pacific regions. The size of the Hispanic populations in these latter areas may contribute to the longer duration of spells, given that Hispanic children appear to have the highest frequency of long spells. Finally, the East North Central and mountain regions join New England in having relatively short durations, but beyond this the regional differences are inconsistent.
With relatively weak differentials in the duration of uninsurance spells by demographic characteristics, we would not expect the distribution of demographic subgroups of uninsured children to vary much by spell duration. Table 15 shows the distribution of demographic subgroups of uninsured children within each grouping of spells by duration. Age composition shows no meaningful variation by spell duration. Race and ethnic composition show very limited variation. Non-Hispanic white children represent a somewhat larger share of 1 to 4 month spells than they do of longer durations while Hispanic children represent a larger share of spells exceeding 12 months than they do of shorter spells. Non-Hispanic black children appear with relatively greater frequency among spells of 5 to 12 months than they do among shorter or longer spells. Children in mother-only families represent 35 percent of 1 to 4 month spells compared to only 21 percent of 13 month or longer spells. Children in two-parent families show the opposite tendency. They represent 69 percent of the longest spells but only 55 percent of the shortest spells. Metropolitan residence continues to show no consistent variation while there are small regional differences. Children from New England and the East North Central states appear with greater frequency among 1 to 4 month spells than among longer spells while children from the Pacific states account for only 18 percent of the 1 to 4 month spells but 28 percent of the 13 month and longer spells.
Like demographic characteristics, socioeconomic characteristics display a much weaker association with spell duration than with the incidence of uninsurance. In Table 16 we see that poverty has a weak, inverse relationship with spell duration. Children from the poorest families tend to experience a higher frequency of very short spells and a lower frequency of long spells than children from less poor or higher income families. We speculate that this may reflect the impact of Medicaid as a source of coverage that is more available to children in very poor families than it is to children in less poor families. The highest incidence of spells lasting 13 months or more occurs
among children between 100 and 300 percent of poverty. Long spells decline in frequency in the top income class. Spell length shows a very inconsistent relationship with parents' employment, with no category being associated unambiguously with long spells or short spells. For example, children with no parent present are most likely to have spells of 1 to 4 months and spells of 13 months or longer. To some extent this inconsistency may be due to the fact that parents' employment is itself a dynamic characteristic. Differentials in the duration of uninsurance may depend on when we measure parents' employment (before, during, or after a spell of uninsurance). In Table 16, as the title indicates, we measure characteristics at the start (that is, in the first month) of a spell, which gives us a uniform measure across spells and increases the likelihood that the characteristics we observe are associated with why the child is uninsured. If we measure socioeconomic characteristics at only one point in time, clearly this is he preferred point to do so. But this is not to say that there is nothing to be learned by examining differentials in duration by income and parents' education measured at another point in time.
The educational attainment of children's parents has a moderately strong, inverse relationship to the duration of spells of uninsurance. This is most evident in the proportion of children with durations of 13 months or longer, which declines monotonically from 33 percent to 12 percent as parents' education rises (excluding the children with no parent present). It shows up as well in the proportion of children with spell durations of 8 months or less, which rises--also monotonically--from 60 percent to 81 percent as parents' education increases.
Socioeconomic differentials are even less evident in the shares of children completing spells of different durations (Table 17). Children in families below 50 percent of poverty represent an increasingly smaller share of spells as duration rises, but there are no other consistent patterns by poverty level. This lack of consistency is true of differences in shares of spells represented by children classified by their parents' employment and parents' education. The fact that we see such weak differences in spell shares by parents' education despite the fairly strong differences in the distribution of spell length by parents' education appears to be due to the small numbers of uninsured children whose parents fall into the lowest and highest levels of education. Children whose parents completed less than 7 years of schooling account for twice as many of the longest spells as they do all shorter spells, but this represents only a five percentage point difference. Similarly, children whose parents completed some graduate school represent only half as many of the longest spells as they do spells of 8 months or less, but this is only a 4 percentage point difference.
3. Duration of New Spells of Medicaid-Eligible Uninsurance
As with all spells of uninsurance, the duration of spells of Medicaid-eligible uninsurance shows similarly weak relationships to the demographic and socioeconomic characteristics of uninsured children. Part of the explanation is the very high proportion of spells that are completed in 4 months or less: 75 percent (Table 18). This leaves little room for significant variation across population subgroups.
Only infants show a pattern that differs to any degree from that of other age groups. Infants are somewhat less likely to have spells of one 1 to 4 months than the other age groups--65 percent versus about 76 percent--but they are no more likely than other groups to have very long spells. Non-Hispanic white children have both the highest frequency of very short spells, at 81 percent, and the lowest frequency of very long spells, at just under 5 percent. Other differences by race and ethnicity are negligible. Children in two-parent or father-only families also have the highest frequency of very short spells, at 77 to 79 percent, and the lowest frequency of very long spells, at 5 percent. The corresponding percentages for children in mother-only families are 70 percent and 9 percent, respectively. Children from no parent families compare at 70 percent and 10 percent.
TABLE 18 (PDF document.)
There is a 10 percentage point difference between metropolitan and non-metropolitan children in the proportion of spells completed in 1 to 4 months. The children to whom the metropolitan/non-metropolitan classification is not applicable differ from the other two groups in the proportion of spells that are completed in 1 to 4 versus 5 to 8 months and in the proportion running 13 months or longer, but the uncertainty about who this group represents makes it unclear what this is telling us.
Because the number of children experiencing spells of Medicaid-eligible uninsurance is much smaller than the number who experience any spells of uninsurance, sampling error makes a greater contribution to the differentials in these tables than the earlier tables, and this is nowhere more
evident than in the regional differences. The West North Central and Pacific regions have the highest percentages of very short spells, at 80 to 81 percent, and among the lowest frequencies of very short spells, at 0 to 3 percent. The South Atlantic and West South Central regions are similar. Unlike all of the earlier tables, the New England region shows nothing distinctive in its patterns. In general, the regional patterns in Table 18 deviate so much from what we have seen earlier that we should be cautious in attaching much credence to them.
Table 19 reports shares of spells within each duration group by demographic characteristics. In general, we see little variation across the columns. Non-Hispanic white children account for 56 percent of the shortest spells compared to 39 percent of the longest spells while non-Hispanic black children represent 26 percent of the longest spells compared to 17 percent of the shortest spells. Similarly, children in two-parent families account for 66 percent of the shortest durations compared to 51 percent of the longest durations while children from mother-only families account for 26 percent of the shortest durations but 40 percent of the longest. A comparison with Table 8 indicates that these findings are consistent with the relative frequency of Medicaid eligibility among uninsured children by race and ethnicity and family composition although they appear to run counter to the
TABLE 19 (PDF document.)
differentials that we find in Medicaid participation (see Table 3). Clearly, a high participation rate in Medicaid does not imply that children leave spells of Medicaid-eligible uninsurance relatively quickly.
Metropolitan area children account for a bigger share of long spells than short spells at 76 percent versus 68 percent while non-metropolitan area children account for 31 percent of the shortest spells and 22 percent of the longest. Regional differences again show the effects of particularly large sampling error. Every region displays fairly wide variation in its shares of spells of different durations, but these differences do not yield a clear pattern.
Differences in the duration of spells of Medicaid-eligible uninsurance by socioeconomic characteristics are much weaker than the differences we reported in the duration of all spells of uninsurance (compare Table 20 with Table 16). There are negligible differences across the first three poverty classes, where sample sizes are largest, and the two highest poverty classes are not substantially different, given their very small numbers.(10) Children with working parents, whether full time or part time, tend to have a greater frequency of short spells than children with no working parents or no parents present and a somewhat lower frequency of very long spells. Differences in spell duration by parents' education are very inconsistent, suggesting that what differences we do see are strongly affected by sampling error. Children whose parents attended graduate school clearly have the shortest spells, with 82 percent being 1 to 4 months in length and only 1 percent exceeding 8 months. Children whose parents completed less than 7 years of schooling have the longest spells, with 60 percent being 1 to 4 months and 23 percent 9 months or longer. But children whose parents completed college (and went no farther) look more like children whose parents completed only 7 to 11 years of schooling than they resemble children whose parents attended but did not complete college or completed college and went on to graduate school.
TABLE 20 (PDF document.)
There are somewhat greater differences in the shares of spells of different durations by socioeconomic characteristics (Table 21) than by demographic characteristics (Table 19). Children in families below 50 percent of poverty account for smaller shares of spells 9 months or longer than they do of spells completed in 8 months or less, but patterns at higher family income levels are less clear. Children in families above 100 percent of poverty represent somewhat larger shares of the spells lasting 9 months or longer than they do of spells ending in 8 months or less. Children without working parents account for larger shares of spells 9 months or longer than they do of shorter spells while children with at least one parent working full time show the reverse pattern, but even here sampling error is evident in the volatility of the shares. Finally, children whose parents completed 7 to 11 years of schooling account for 33 percent of the longest durations compared to no more than 22 percent of any class of shorter durations whereas children whose parents attended but did not complete college represent 14 percent of the longest durations compared to 20 percent of the shortest.
4. Duration of Active Spells of Uninsurance
Earlier in this section we examined demographic and socioeconomic differentials in the lengths of completed spells of uninsurance. Spells that are active at a point in time have a different distribution--both currently and, especially, when completed--than do spells that started during a year. Furthermore, because active spells are generally not complete (some will end in that month, but most will continue), the relationships beween their duration and various characteristics of the uninsured differ from what we would observe with completed spells. More specifically, the shorter the active duration, the greater the proportion of these spells that will end up with longer completed durations. In terms of their characteristics, children who are in short active spells will look much more like children with completed spells of all durations than they do like children with completed spells of short duration. We would be much less interested in the characteristics of uninsured children by how long they have been uninsured were it not for the fact that state CHIP plans frequently limit coverage to children who have already accumulated some specified number of months of uninsurance. Table 22 compares the distributions of active spell lengths and completed spells by three characteristics: the child's age, family poverty level, and parents' employment.
TABLE 21 (PDF document.)
Active spells are limited to those of children who were uninsured in September 1993 while the completed spells represent all spells that were started in FY93.
The active spells include a higher proportion of very long spells than do the completed spells; 46 percent of the children uninsured in September 1993 had already been uninsured for 13 months or more whereas only 20 percent of all the spells that started in FY93 (which may include multiple spells by the same individuals) extended beyond 12 months.
Infants have much shorter active spells than completed spells because spell lengths are bounded by their very limited life spans. We see no infants with spells exceeding 12 months, of course, because no infant, by definition, has been alive for more than 12 months. By contrast, there is no limit to the completed spell length of children who were infants when their spells began, and infants have nearly the same proportion of completed new spells exceeding 12 months as older children: 17 percent compared to 18 to 22 pecent for older children). The proportion of older children whose active spells exceed 12 months rises 20 percentage points--from 36 percent to 56 percent--between ages 1 to 5 and ages 11 to 15 but then drops slightly, to 49 percent, in the next age group. The proportion of new spells exceeding 12 months in length shows no increase with age after infancy, however. It is not clear why we should see an age difference in the current duration of active spells when there is no age difference in the completed durations of new spells. Active spells are in some sense sampled from new spells but with a probability proportional to (completed) length. It must be that the age differences that we see among active spells are due to age differences that do not appear in new spells until durations well beyond 13 months.
TABLE 22 (PDF document.)
There are differences in the durations of both completed spells and active spells by poverty level, but the differences among active spells are weaker. The proportion of children with completed spells of 1 to 4 months declines from 63 percent to 51 percent over the first three income classes while the change among active spells is only 6 percentage points. Similarly, the proportion of new spells exceeding 12 months in length rises from 14 percent to 23 percent over the first three income classes whereas the increase among active spells is only 5 percentage points. Both distributions show an upturn in the relative frequency of short spells as income rises to more than 300 percent of poverty.
Parents' employment shows little relationship to the duration of completed spells. Children with no parent present have a somewhat higher frequency of short spells than other children, but there are no other consistent differences. Among children with active spells, those with no parents present have fewer spells exceeding 12 months than other children but comparable numbers of spells exceeding 8 months. Children with no working parents are somewhat more likely to have been uninsured for less than 5 months than children with working parents, but again the differences are modest.
Differences in the shares of active spells by demographic and socioeconomic groups are of interest because of what they tell us about the composition of children who may be eligible for state CHIP coverage. As we have pointed out, infants by definition cannot be uninsured for more than 12 months. States that would like to cover infants but choose to limit their coverage to children who have been uninsured for 12 months or more or even 6 months or more will have to consider defining eligibility among infants on some basis other than the length of their own spells of uninsurance, or infants will receive disproportionately little coverage. In Table 23 we see that while infants account for 7 percent of the spells of uninsurance that started 1 to 4 months earlier, they account for only 3 percent of the spells that started 5 to 8 months earlier and closer to 2 percent of the spells that started 9 to 12 months earlier. Shares of active spells that started more than 12 month earlier rise with age.
Children 1 to 5 account for 18 percent of such spells while children 11 to 15 account for 35 percent. There is little difference, however, in the age group shares of spells lasting 9 months or longer.
Both active spells and completed spells show the share of children in families between 100 and 200 percent of poverty rising with duration. This group accounts for 36 percent of the active spells that started 1 to 4 months earlier compared to 44 percent of the spells that started more than 12 months earlier. Completed spells show the share of children in this group rising from 34 percent to 41 percent. Other poverty classes show less variation in their shares of children with different durations of uninsurance.
Shares of the uninsured by parents' employment show minimal variation by parents' employment. For active spells, no employment group's share varies by more than 4 percentage points. There is somewhat greater variation among completed spells, but no group shows consistent growth or decline in its share of all spells as duration increases.
On the whole, then, spells active at a point in time show no greater evidence of demographic or socioeconomic differentials in duration than new spells that started over the course of a year.
TABLE 23 (PDF document.)
When we measure age at a point in time but measure behavior over a period of time, some of the behavior that we observe may have happened at an earlier or later age. If the behavior in question is affected by crossing particular age boundaries--as is Medicaid participation--then the relationship that we observe between age and the measured behavior may be attenuated by defining age at a point in time. Tables 24 through 27 are based on a different approach to assigning age to behavior measured over time. These tables examine the experience of uninsurance, Medicaid participation, and both Medicaid-eligible and Medicaid-ineligible periods of uninsurance through the 12 months that a child spends at a given single year of age. To create these tables, we identified children by their age in September 1993 and then looked forward and backward to determine the 12-month period that each child spent at the defined age.(11) We then counted the number of months over this 12-month period that each child was observed in each of four statuses: uninsured (Table 24), uninsured but eligible for Medicaid (Table 25), uninsured but not eligible for Medicaid (Table 26), and enrolled in Medicaid (Table 27). For each of these four statuses we tabulated by single year of age:
TABLE 24 (PDF document.)
TABLE 25 (PDF document.)
TABLE 26 (PDF document.)
TABLE 27 (PDF document.)
These tabulations give us a clearer picture of the age-specific patterns of uninsurance and Medicaid participation than we obtain when we classify children's behavior over the course of a year by their age at the beginning or end of that year.
1. Uninsurance
In Table 24 we see that among infants in September 1993, 16.4 percent were ever uninsured during their infancy. Among those who were ever uninsured the average number of months without insurance was 4.87, and 9.6 percent of these children were uninsured for all 12 months. The proportion of all infants who were uninsured for the entire year of their infancy was 1.6 percent. From the average duration of uninsurance we could calculate the probability that an infant who was ever uninsured during the year was uninsured at a point in time. This is simply the average number of months uninsured divided by 12. Multiplying this probability (.406) by the percentage ever uninsured during the year gives us the average percentage of all infants--6.7 percent--who would have been uninsured at a point in time. For 18-year-olds this same calculation would yield 17.3 percent as the proportion likely to have been uninsured at a point in time.
The percentage of children ever uninsured during the year shows a progressive rise from age 0 to 18, reaching 26.9 percent among 18-year-olds. The average duration of uninsurance rises as well but not as steeply as the percentage of ever-uninsured children who were uninsured for the entire year. This latter approaches 40 percent among the oldest children whereas it is less than 10 percent among infants and just over 20 percent in the next higher ages. (These estimates show marked variability among neighboring ages, which we attribute primarily to sampling error.) The percentage of all children uninsured for 12 months rises even more dramatically because it is the product of two percentages that rise with age: the percentage ever uninsured and the percentage among these who were uninsured the entire year. From a low of 1.6 percent among infants and 4.0 percent among children one year of age the proportion uninsured for all 12 months rises to just over 10 percent among children age 18.
Turning to Medicaid eligibility within spells of uninsurance, we find very different patterns by age than we do for uninsurance generally. In Table 25 we see that just under 11 percent of all children were ever Medicaid-eligible and uninsured during the year. This fraction is 13 percent among infants, and it remains at about that level until age 10, when it drops to between 7 and 8 percent. Eligibility under the poverty-related criteria did not extend to children who were age 10 in September 1993; children 10 and older at that point could qualify for Medicaid through a number of other channels, but the unavailability of coverage through the poverty-related criteria limited eligibility to barely more than half the number who were eligible at younger ages.
That the probability of being both uninsured and Medicaid-eligible varies so little by age can be explained by the fact that Medicaid eligibility rates decline with age while the probability of being uninsured rises. With these two rates moving in opposite directions their product tends to remain nearly constant. By the same reasoning, however, we would expect the probability of being uninsured and not eligible for Medicaid to rise sharply with age. Both the probability of being uninsured and the probability of being ineligible for Medicaid increase with age. Hence the age pattern will be amplified when these two statuses are combined.
The average number of months that children were uninsured and Medicaid-eligible did not vary by age, it appears.(12) Children remained in this state for an average of nearly five months at every year of age. We suggest that the uniformity of this experience by age may be due to the fact that, unlike spells of uninsurance without Medicaid eligibility, parents could choose to end a spell of Medicaid-eligible uninsurance by enrolling the child in Medicaid. There is no obvious reason why the age of the child should affect how long parents choose to wait, except perhaps for infants, although spells for infants were no shorter than spells for older children.
The proportion of ever-uninsured (and Medicaid-eligible) children who were uninsured for all 12 months shows very considerable sampling error. This proportion seems to rise with age whereas the percentage of all children who were uninsured and Medicaid-eligible for all 12 months remains steady or even falls.(13)
Periods of uninsurance without Medicaid eligibility occurred with a frequency between that of Medicaid-eligible uninsurance and all uninsurance (Table 26). Because uninsured children below age 10 were nearly twice as likely to be eligible for Medicaid as children 10 and older, the proportion of children who were ever uninsured and not eligible for Medicaid during the year rises more sharply with age than the percentage who were simply ever uninsured. From 6 percent among infants the percentage ever uninsured and not Medicaid-eligible rises to 24 percent among 18-year-olds. Mean durations in this state rose by two to three months over the range of ages, with an average duration of 6.2 months, while the percentage of those remaining in the state for 12 months grew from the low single digits to over 30 percent, with an average of 20 percent. Among all children, the percentage who were uninsured without Medicaid eligibility for the entire year rose from .3 percent among infants to nearly 8 percent among 18-year-olds.
2. Medicaid Participation
Table 27 presents statistics on reported Medicaid coverage or participation by single year of age. Among all children, 24.2 percent were ever covered by Medicaid during the year. Those who were ever covered were covered for an average of over 9 months, and 55.1 percent were covered for the entire year. Among all children the percentage covered for all 12 months was 13.3 percent.
There is a steep age gradient in Medicaid participation, but it does not show the clear break between ages 9 and 10 that we see in the tabulations of simulated eligibility. Participation is highest among infants, with 39 percent ever covered during the year. Infants with coverage are enrolled for an average of more than 10 months of their infancy, with about 64 percent of this group or 25 percent of all infants covered for the entire year.(14) The percentage of children ever covered during the year drops to 35 percent for ages one and two and then falls to between 28 and 30 percent through age six. From about 25 percent at age nine the proportion ever covered during the year descends to 21 percent at age 10, then declines gradually to 19 percent by age 14, drops to 15 percent by age 16, rises to 19 percent, and then plunges to 12 percent at age 18.
Among children ever covered by Medicaid, both the average number of months of coverage during the year and the percentage covered for all 12 months decline with age. The average number of months of coverage drops from 10.3 months for infants to 8 months by age 18. The percentage of children covered for all 12 months falls from 64 percent among infants to 33 percent at age 18. Among all children, the percentage covered by Medicaid for all 12 months of the year shows an even more dramatic decline, from 25 percent among infants to just 4 percent among 18-year-olds.
This report has used data from the 1992 panel of the SIPP to examine demographic and socioeconomic differentials in the patterns of health insurance coverage among children under 19. Health insurance coverage among children varies by nearly every demographic and socioeconomic characteristic that we examined. Most of the differentials that we observe in the type of insurance coverage and whether there is any coverage at all are moderately strong to very strong. For example, Hispanic children are more than two-and-a-half times as likely to be uninsured as white non-Hispanic children, and black children are four times as likely as white children to be covered by Medicaid.
Because of Medicaid, coverage patterns are not unidimensional. Groups with low rates of employer-sponsored coverage do not necessarily have high rates of uninsurance. High rates of Medicaid coverage can appear among groups with high uninsurance or moderately low rates of uninsurance. Nevertheless, the strongest differentials by far are those associated with parents' education, and these differentials are strikingly unidimensional.
We estimate that one-third of the children uninsured in September 1994 were eligible for Medicaid, based on a Medicaid eligibility simulation that takes account of most but not all of the ways that children may become eligible. There are sizable differentials in Medicaid eligibility across most of the demographic and socioeconomic characteristics that we examined. Differentials in Medicaid eligibility among the uninsured are strongest by family poverty level, followed by the child's age and parents' employment.
Over all eligible children, the estimated Medicaid participation rate is 65 percent. This participation rate does not reflect any adjustment for the underreporting of Medicaid participation in the SIPP, which in 1994 was on the order of 24 percent. If we exclude from the eligible nonparticipants those who were covered by insurance other than Medicaid, however, the participation rate rises to 79 percent. Differentials by demographic characteristics are largely eliminated when the participation rate is defined in this alternative way, and socioeconomic differentials are greatly reduced.
Generally, differentials in the proportion of children ever uninsured during the year are similar in form to differentials in the percentage uninsured at a point in time. Only for race and ethnicity and region did we find that groups were arrayed differently with respect to their likelihood of being uninsured. Distributions of the uninsured by demographic or socioeconomic characteristics differ very little between the point-in-time and annual-ever measures.
The duration of new spells of uninsurance shows little variation by demographic characteristics or parents' employment. There is modest variation in spell length by poverty level and moderately strong variation by parents' education, however. Spell lengths decline with parents' education.
The current duration of active, or ongoing, spells--that is, how long uninsured children have been without coverage--is of interest recently because a number of states are planning to limit eligibility under their CHIP initiatives to children who have been uninsured for some minimum number of months--as many as 12 months. It appears that such restrictions favor the group that is the principal target of CHIP: children between 100 and 200 percent of poverty. This is the one subgroup whose share of the uninsured clearly increases with duration.
We also examined the insurance coverage of children's parents. Between 18 and 21 percent of uninsured children have an insured parent. This suggests that a nontrivial number of parents may be choosing individual coverage but not family coverage--perhaps because family coverage is not offered or is perceived as too expensive. Measurement error may account for part of this finding as well. Parents may report their own coverage but overlook the coverage that extends to their children. This seems particularly likely for the 7 percent of uninsured children who report that a parent is covered by Medicaid. Among children who are themselves reported to be covered by Medicaid, 10 percent have an uninsured parent. This latter is consistent with the child-only eligibility created by the poverty-related expansions of Medicaid in the late 1980s and 1990s.
Finally, to provide detailed information on life cycle patterns of coverage among children, we estimated the frequency with which children experienced periods of uninsurance, with and without Medicaid-eligibility, during each single year of age. We also estimated the frequency with which children experienced periods of reported Medicaid enrollment. Both the probability of being uninsured and the average number of months uninsured among those with any months of uninsurance increase with age. The likelihood of being uninsured for all 12 months shows a particularly strong relationship with age. The probability of being uninsured and eligible for Medicaid has little relationship with age, however, being relatively constant from infancy through age 9 and then dropping a few percentage points to a level that remains essentially fixed through age 18. The average number of months of Medicaid-eligible uninsurance is relatively constant across the entire age range as well. By contrast, uninsurance without Medicaid eligibility is even more strongly associated with age than uninsurance in general. The estimates of Medicaid participation by single year of age make clear the prominent role of Medicaid as a source of coverage among infants and the increasingly smaller role that Medicaid plays as children grow older. As more children become eligible for the federally mandated coverage for children below the poverty line who were born after September 30, 1983, however, these age differentials will gradually diminish.
1. Each year the upper age limit for federally mandated coverage rises by one year, in effect.
2. Foster children may be covered by Medicaid and, depending on whose household they appear in the SIPP, could be reported as having no parents in the household.
3. The SIPP instrument includes questions on insurance coverage provided to household members by persons outside the survey household, but it is not difficult to imagine that such coverage is reported less completely than coverage provided by parents or other adults in the household.
4. The poverty thresholds, which are provided on the SIPP file, are the same thresholds that the Census Bureau used to calculate the official estimates of children in poverty in 1994.
5. Monthly poverty rates run higher than annual poverty rates, generally, but annual poverty rates obtained from the SIPP by aggregating monthly income and poverty thresholds for individuals tend to run lower than the official poverty rates estimated from the Current Population Survey (CPS). The difference in the annual estimates can be attributed in large part to the SIPP's more accurate attribution of income to family members actually present each month.
6. These estimates of Medicaid eligibility are based on a detailed simulation described in Technical Appendix A. The simulation uses monthly income (in this case for September 1994) and other characteristics--such as the child's age and family composition--measured in the same month, along with state-specific eligibility criteria. The use of monthly data yields a more accurate simulation of the actual Medicaid eligibility determination than does the use of annual data or characteristics measured only once during a year. Other things being equal, monthly income would yield more eligibles than annual income, but there are a number of factors that confound the comparison of our monthly simulation with what other researchers have done with annual data. Finally, our simulation does not include the spend down features of the medically needy program, because the SIPP does not collect medical expenditure data. Therefore we know that we understate eligibility for Medicaid. We are not aware of any other simulations that are comparable to ours in other respects but include this feature of eligibility.
7. We exclude children who were reported to be participating in Medicaid but were simulated to be ineligible. We do so because many of the simulated ineligible participants may have been eligible for Medicaid under provisions that we did not simulate. To include just the participants would be equivalent to assuming that there were no nonparticipants among this additional group of eligibles when the participation rates that we observe for children eligible under related provisions suggest that there may have been at least twice as many nonparticipants as participants.
8. If some of these children were actually enrolled in Medicaid and their coverage had simply been misreported, we would want to include them in both the numerator and denominator of the participation rate. If all of these children were actually enrolled in Medicaid, including them in this way would yield a Medicaid participation rate of nearly 83 percent.
9. It is plausible, as well, that some number of spells that lasted fewer than four months were not reported at all. On balance, we suspect that spells of 1 to 4 months in length are still overstated, but the potential exclusion of some spells altogether has implications for the estimated number of children who were uninsured at a point in time or ever in a year. In fact, all misreporting of durations may affect estimates of incidence as well..
10. Fewer than 5 percent of the uninsured children in families above 200 percent of poverty were simulated as Medicaid eligible.
11. For example, a child who was observed at age 1 in September 1993 could have turned 1 in any of the months from October 1992 through September 1993 and ended that year of age in any of the months from September 1993 through August 1994.
12. We attribute the variability that we see to sampling error.
13. Recall that the proportion eligible for all 12 months among all children is the product of the proportion eligible all 12 months among those ever eligible, which rises with age, and the proportion who are ever eligible, which declines with age.
14. The 39 percent of infants whom we find to have ever participated in Medicaid during their infancy and their average of 10.3 months of coverage imply that 33.5 percent of infants would have been covered by Medicaid at any point in time. This compares quite closely to the 34.2 percent who were reported as covered in September 1993 (Table B.3).