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Explaining the rise in antebellum pauperism, 1850–1860: New evidence
The Quarterly Review of Economics and Fmance, Vol. 37, No. 2, Summer 1997, pages 403-417 Copyright 0 1997 ‘Ihutees of the University of Illinois All rights of reproduction in any form reserved ISSN 1062-9769
Focus
Explaining the Rise in Antebellum Pauperism, 1850-1860: New Evidence
L. LYNNE KIESLING College of William & Mary
ROBERT
A. MARGO
Vanderbilt University
Between 1850 and 1860 the total ‘$auper rate”A”_
number of in&viduuls receiving public assistance per 1,000 population-increased j&n 5.8 to 10.2. We explore the determinants of the rise in antebellum pauperism using previously enexploited archival data. Changing labor market conditions, urbanization, and immigration led to a marked increase in the Lmund for public assistance. Antebellum taxpayers, however, were unwilling to maintain the generosity of relief at existing levels in the face of the rise in demand.
The 1850s were a decade of contrasting economic fortunes in the United States. According to Robert Gallman’s unpublished estimates, real per capita income increased by 24.5% between 1850 and 1860.’ Frontier states grew rapidly as population migrated westward in pursuit of economic gain. Railroad mileage increased by 360% over the decade, facilitating an enormous expansion in internal trade (U.S. Secretary of the Interior 1866, p. 331; Taylor, 1951). Against the backdrop of economic growth, the 1850s also witnessed a dramatic rise in the incidence of public “pauperism.” According to the 1850 census, the total pauper rate-the total number receiving poor relief from public funds during the year, per thousand persons- was 5.8/1,000 in 1850. B 1860 the total pauper rate had increased to 10.2/1,000, or by 76% in ten years. Y Although previous work has recognized the increase in pauperism during the 185Os, there is no consensus on the relative economic significance of different factors (Hannon, 198413, 1996; Trattner, 1994). This paper uses previously untapped archival evidence, in conjunction with published census data, to explore the determinants of public pauperism during the 1850s. The data, which are derived from the manuscript censuses of social statistics of 1850 and 1860, are sufficiently rich to permit us to estimate an econometric model of pau-
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per rates at the county level. Our principal findings are that changing labor market conditions, urbanization, and immigration led to a marked increase in the demand for public assistance between 1850 and 1860. However, antebellum taxpayers were apparently unwilling to maintain the existing level of financial generosity of public assistance in the face of increased demand.
I.
BACKGROUND
DISCUSSION
Since the New Deal the provision of public welfare in the United States has been the primary responsibility of the federal government. Recent reforms, however, are in the process of shifting that responsibility back to state and local governments. In decentralizing welfare policy, current reform efforts are gradually returning the system to its historical roots-that is, to a system in which “local”, as opposed to “national”, governments paid for, and implemented, relief policy. Antebellum poor relief was a diverse affair. This diversity reflects the local nature of the English poor law, from which American poor relief policy was derived during the colonial period (Trattner, 1994). American colonial statutes preserved the family responsibility for the needy, with public relief available for those unable to care for themselves, or whose families were unable to do so. During the antebellum period local governments (primarily at the county level) were fiscally responsible for providing relief. Recipients generally had to meet some form of residency requirement. Public provision of relief included the poorhouse (indoor relief), and cash or in-kind transfers (outdoor relief). Public officials favored poorhouses, on the belief that outdoor relief had more severe disincentive effects on work effort (Trattner, 1994). Contemporary observers recognized that pauperism was increasing in the 1850s. Although many attributed rising pauperism to moral failings on the part of the poor, the traditional view among historians places the blame for the rise in pauperism squarely on urbanization, industrialization, and immigration (Hannon, 1984b, p. 1018). Urbanization was at fault because the urban poor allegedly lacked access to the social networks (family and friends) that traditionally sustained the rural poor through difficult times. The displacement of the artisanal shop with the factory system and with “outwork” reduced the skill requirements of industrial labor and with it, household income, putting industrial workers at greater risk of needing occasional public assistance (Wilentz, 1984). Immigration was a culprit for reasons similar to industrialization-the foreign born had lower incomes than the native born and, therefore, more prone to seek relief when faced with an adverse economic shock. Recent research has provided a more nuanced view of the possible causes of rising antebellum pauperism (Katz, 1983; Hannon, 198413, 1996). Labor demand, rural and urban, was highly seasonal during the antebellum period. In rural areas, however, under-employment was more readily disguised, and farm households could generally meet their consumption needs through home pro-
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duction. But, with the spread of wage labor (via industrialization and urbanization), workers became more dependent on market-generated incomes. Unable to engage in home production during bouts of seasonal (or cyclical) unemployment, the urban poor increasingly turned to public assistance. Lacking the savings required to smooth consumption, low-wage workers were especially vulnerable when, as in the 185Os, the price of food relative to unskilled wages rose sharply (Komlos, 1987, p. 915; Margo, 1997, ch. 2). Analysis of administrative data for New York State by Joan Hannon (1984b) provides some support for this nuanced view of the antebellum welfare explosion. Between the late 1840s and late 1850s pauper rates in New York state rose by nearly 60%. Yet residency rates in poorhouses remained roughly constant, implying that the vast majority of the additional paupers were receiving outdoor relief for relatively brief periods of time (Hannon, 198413, pp. 1012, 1018).3 Although Hannon’s data did not permit her to examine directly the relationship between low wages and pauperism, she was able to show that county-level pauper rates were negatively correlated with the extent of household production, and positively with urbanization and the percent foreign born (Hannon, 1984b, p. 1026).
II.
ARCHIVAL
DATA ON POOR RELIEF
Although various states produced statistics on the incidence and cost of public poor relief beginning in the early 18OOs, the first attempt to produce comparable national statistics on poor relief occurred in 1850. As part of the enumeration effort in 1850 and again in 1860, census marshals were instructed to collect information for minor civil divisions (MCDs) on the total number of recipients of poor relief (native and foreign); the number receiving poor relief on June 1 (native and foreign); and total expenditures on poor relief. At the time the census was compiled, the data were aggregated to the state level, and state totals were reported in the published volumes of the 1850 and 1860 censuses. Economic historians have used these state totals in previous work on antebellum poor relief (Lebergott, 1976). The original manuscripts of the social statistics survive for some states, however, and microfilm copies are available at the National Archives or from various state archives. For the purposes of this paper we obtained and computerized data from microfilms for eight states, two from each census region: Pennsylvania and Massachusetts (Northeast); Michigan and Iowa (Midwest); North Carolina and Virginia (South Atlantic); and Kentucky and Tennessee (South Central). The selection of states was dictated by the cost of data retrieval, balanced against the desire to have a regionally dispersed sample of counties. The instructions to census enumerators specified that social statistics were to be collected for each MCD (or townships, in the case of New England) “as far as practicable” (DeBow, 1854, pp. xxiv). Inspection of the microfilms suggest that
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this goal was broadly achieved, in the sense that virtually all counties reported data for at least one MCD. Sometimes the reported MCD names correspond to known geographic entities (for example, wards in Philadelphia) but in other cases they do not (for example, a census marshal might refer to an MCD as “my division”). Because of this problem (along with illegibility of some MCD names) we have chosen to aggregate all data to the county level. Other than commenting that information was “not to be ascertained entirely by personal inquiry of individuals, but in part from public records and reports, and public offices” the census was silent on the exact procedures marshals were to follow in collecting the information on poor relief (Debow, 1854, p. xxiv).4 Unfortunately, none of the microfilms contain marginalia on data collection procedures. However, “public records” probably provided most or all of the relevant figures, on the assumption that individuals chosen randomly by the census enumerators would presumably be poorly informed about the number of persons on relief or the amount spent on relief.5 The problems with the data are numerous. No information is available on the type of relief (outdoor versus indoor), or on age or gender characteristics, although Hannon’s (198413) analysis of New York administrative records suggests that the increase in pauperism in the 1850s was not caused by demographic shifts.6 Direct information on the average duration of relief (for example, in days or months) was not reported. Lebergott (1976), however, suggested that the number of recipients on June 1 was a good proxy for the “long-term” relief population.’ The implication of Lebergott’s suggestion is that the remaining recipients (Total recipients - June 1 recipients) can be thought of as “shortterm.” The appendix to the paper develops a method to infer the average duration of relief from the census distinction between short and long-term recipients, and our estimates of average duration are consistent with Lebergott’s interpretation-the predicted duration of relief was considerably shorter for “short-term” recipients (as Lebergott defined them) than “long-term” recipients (see Appendix Table 3). This study uses our estimates of duration to construct “lull-time equivalent” (FTE) measures of pauperism. In addition, many MCD’s, particularly in frontier areas (for example, various counties in Iowa in 1850) simply left the relevant columns blank, which we have interpreted as true zeros. Upon aggregating to the county level, nonreporting is reduced but not eliminated. However, once the data are weighted by countylevel population counts, the economic significance of “zeros” is greatly reduced. All the numerical estimates reported in this paper are weighted by relevant county-level populations. Sample statistics on pauper rates are reported in Table 1. Consistent with the direction of change revealed by the published census aggregates, the total pauper rate in the sample increased between 1850 and 1860 from 4.6/1000 to 7.8/1000, although the levels of pauperism in the sample are below (in both years) the aggregate levels reported in the published census. Consistent with Hannon’s (1984b) findings for New York State, the rise in the total pauper rate
EXPLAINING
Table1. Short-term Long-term Total FI’E Total
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Sample Statistics: Paupers per 1,000 Population 1850
1860
2.5 2.1 4.6 2.2
5.4 2.4 7.9 3.2
Soum: Eight-state sample fmm manuscript censuses of social statistics.
between 1850 and 1860 was dominated by the rise in short-term relief. Because the duration of relief among short-term recipients was shorter than among longrun recipients, the increase in ETE pauperism between 1850 and 1860 was less than the increase in the total pauper rate. The final row of Table 1 computes FTE pauper rates for the sample, using the estimated durations reported in the appendix as weights (see the appendix). In FTE terms, the antebellum welfare explosion was less dramatic-instead of a 70% (= 7.8/4.5 - 1) increase, the FTE rate rose by 45% (= 3.2/2.2 - I).’
III.
A MODEL OF ANTEBELLUM
POOR REFIEF
To motivate our empirical analysis this section sketches a simple model of public poor relief. The model consists of two functions: a “demand-for-relief’ equation and a “supply-of-relief’ equation.g Both demand and supply are determined at the local level, which for the purposes of our analysis is taken to be the county.” The demand for public poor relief is p = d(X, e), where p is the incidence of public pauperism (for example, the number of paupers per 1,000 population); X is a set of exogenous variables that shift the demand for relief; and e is the “generosity” of relief, assumed to be measured by expenditures per recipient. Following the previous discussion, both p and e are expressed in FTE units. By “demand” we mean that pauperism and generosity are positively related-holding other factors (X) constant, the more generous is relief, the greater the number of persons seeking relief (d, > 0). The supply-of-relief equation is p = ~(2, e), where 2 is a set of variables that shift the supply function. We think of supply as the outcome of an maximization process in which antebellum taxpayers agree to levy local taxes (for example, on property) to fund a socially optimal level of public relief. The exact motives for supplying relief are left unspecified, although at some level we presume that taxpayers are altruistic towards the poor- that is, public assistance was a normal good. Because the marginal utility of helping those is need is assumed to be decreasing, however, and because taxpayer resources are finite, there is a negative tradeoff between p and e on the supply side-holding constant 2, taxpayers are less willing to be generous (se < 0) as p increases. * ’ Equilibrium is reached when p and e are consistent with each other-that is, at a given level of e, the number of paupers desiring relief just equals the num-
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ber that taxpayers are willing to support. Note that if various factors change to cause the demand for relief to increase, p will rise but e will fall. Conversely, if taxpayers become more willing to support (for example, as their wealth increases) p will also rise but so will e. For estimation purposes we assume that bothf and g are linear: Demand: p = xa
+ J3e + E/j
Supply: p = zr - Se + E, where a, J3. I-, and 6 are parameters, E’S are random errors. Included in X are four variables: the “real unskilled wage,” defined to be the full-time annual wage of unskilled labor without board, deflated by the cost of board (food); a dummy variable indicating the presence of an urban area in the county of population 10,000 or more; the percent foreign born; and the number of churches per capita.‘* Like Hannon (1984b) we do not have a direct measure of the extent of wage labor. However, if low-wage workers were especially vulnerable to economic distress, as Hannon suggests, we should observe a strong negative relationship between the real unskilled wage and the FTE pauper rate. For the reasons discussed in Section II, the coefficients of the urban dummy and the percent foreign are expected to be positive. Recent work by Kiesling (1996) and Ziliak (1995) suggests that private and public sources of relief were substitutes. The number of churches per capita is included as a proxy for the availability of private relief; its coefficient is expected to be negative. Included in 2 are the urban dummy, churches per capita, and taxable wealth per capita. The urban dummy is included in the supply equation because the cost of providing relief may have differed between urban and rural counties. If, for example, the urban cost was lower, the coefficient of the urban dummy should be positive. Just as the availability of private sources may have lowered the likelihood of those in need to seek public assistance, the willingness to supply public funds may have been reduced. However, a large number of churches per capita may also capture a heightened sense of social responsibility towards the poor-that is, a shift in community tastes for providing more generous relief. Thus, the coefficient of churches per capita in the supply equation is uncertain. The principal variable shifting the supply equation is taxable wealth per capita. If public relief was a normal good, then higher taxable wealth should have increased the supply of relief. Thus the coefficient of taxable wealth should be positive. To maintain consistency with the definition of the real wage variable, wealth per capita and expenditures per recipient are deflated by the cost of board. The wage of unskilled labor, the cost of board, and taxable wealth are drawn from the manuscript censuses of social statistics, which reported these data along
EXPLAINING
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with the those on poor relief. All other data are drawn from the published volumes of the 1850 and 1860 censuses. To complete the specification we include county fixed effects in both the supply and demand equations. We restrict the estimation to a balanced sample of counties-that is, counties that existed in both 1850 and 1860 and which reported complete data. l3 Because we have excluded wealth per capita from the demand equation, and the real wage and the percent foreign from the supply equation, both equations are identified. Thus, after first-differencing, we estimate the equations using weighted two-stage least squares. The weights are the average of the population counts in 1850 and 1860. Sample means and coefficients are presented in Panel A of Table 2. Both regressions, as well as the majority of the coefficients are statistically significant, and the signs of coefficients accord with our prior expectations. On the demand side, labor market conditions evidently mattered, as indicated by the negative coefficient on the real unskilled wage. So, too, did urbanization and immigration: counties that underwent urban expansion during the decade, as well as immigration, witnessed an expansion in the demand for poor relief. Counties with a greater density of churches appear to have had lower demand for public relief, consistent with the belief that public and private assistance were substitutes. On the supply side, there is little evidence that the cost of supplying poor relief was lower in urbanized counties, or that the presence of a large number of churches per capita significantly altered the supply of public assistance. Richer counties, however, clearly were more willing and able to provide public relief, as evidenced by the positive and significant coefficient of per capita wealth in the supply equation. We also find strong evidence, on both the demand and supply side, of a significant relationship between the incidence of pauperism and of the generosity of relief. As hypothesized in our model, pauper rates were a positive function of expenditures per recipient on the demand side, and a negative function of expenditures on the supply side. The supply equation was much steeper, in relative terms, than the demand equation-the number of paupers seeking relief was quite responsive to the generosity of relief, while local governments imposed a sharp negative tradeoff between pauperism and generosity. In Panel B of Table 2 we report the results of a decomposition analysis, in which we use the sample mean values of the exogenous variables to compute the mean values of the supply and demand shift terms. The mean value of the demand shift term was 4.4, which implies that the demand for relief rose, in FTE terms, by 4.4 per 1,000 between 1850 and 1860. However, the actual increase in the FTE pauper rate in the sample counties was 1.67. Although labor market conditions, urbanization, and immigration served to substantially raise the demand for relief in the 185Os, the steepness of the negative tradeoff between pauperism and generosity on the supply side limited the overall increase in pauperism. l4 Antebellum taxpayers were apparently unwilling to
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Demand-Supply Model of Antebellum Pauperism Sample Mean
A. Weighted TSLS Fixed Effects First Difference of:
Estimates,
Real Unskilled Wage
-0.065
Percent Foreign
0.017
Churches/Population
0.083
Urban County
0.068
Real Wealth/Population
0.083
per FTE Pauper
F-Statistic Mean value of dependent Number of Counties
SUPPlY
1.581 (6.040) -20.389) (4.53 1) 52.158 (4.025) -1.943 (3.321) 12.114 (4.227)
0.745 (3.079)
First Differences
Constant
Real Expenditures
Demand
variable
-0.103
(4.552) -3.390 (2.582) 5.69
1.09 426 Demand
B. Decomposition of Demand and Supply Shifts Predicted Shift, 1850-60 Due to: Real Wage Percent Foreign Urbanization Churches per Capita Wealth per Capita
32.549 (4.349) 4.58
-0.320 (0.678) -1.446 (1.397) 1.554
4.44 1.35 0.89 0.82 -0.20
SUPPlY 0.75
-0.10 -0.03 0.13
Notes: 1.
Dependent variable is first difference (1860-1850) of FTE pauper rate (number of FTE paupers per 1,000 population). Pauper rate and expenditures per pauper are endogenous variables. Absolute value oftstatistics in parentheses. Weight is average of 1850 and 1860 population count. Source: eight-state sample from manuscript censuses of social statistics. 2. Predicted Shift equals sum of mean values of exogenous variables times regression coefficients (from Panel A); see text.
maintain the generosity of poor relief at 1850s levels, on average, when faced with the sharply increased demand for public assistance between 1850 and 1860. IV.
CONCLUDING
REMARKS
Despite rising per capita income, the fraction of persons receiving public assistance in the United States rose substantially between 1970 and 1990. Although some scholars attribute the post-1970 rise in welfare incidence to overly generous public assistance, others point to labor market factors, such as immigration and to decreased relative demand for less-educated workers.t5 These modern
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concerns echo those of the lS5Os, which witnessed one of the first. welfare explosions in American history. Labor market conditions, immigration, and structural change led to a marked increase in the demand for public assistance between 1850 and 1860. Antebellum taxpayers were apparently willing to help those in genuine need, but their generosity was strained by the rise in demand and, cons~q~endy, the supply of relief did not keep pace. The results of the paper could be extended in two directions. First, it would be valuable to expand the sample to include additional statesi Second, it would be useful to estimate a similar econometric model for a later (but pre-New Deal) period (if suitable data coulcl be found) to see if the magnitudes of the tradeoffs between pauperism and expenditures on the supply and demand side remained stable over time.
APPENDIX Estimating the Average
This duration
appendix of relief.
Duration of Relief
develops a method for using the census data to estimate Our estimates are derived from the following equation:
the
where
ed= Dg i j Sq i j
= = = = = =
expenditures per recipient per day; average duration of relief, in days, L1‘ (long-term), ST (short term), ~(ative), ~(forei~~); share of relief type in total recipients, LT, ST, N, F.
The data contain no direct information on cd. In place of direct infor~~ation, we make the assumption that ed = ~*(we~kly cost of board/i’). 7hat is, expenditures per day were a constant fraction of the daily cost of food. Because the data on board refer to the cost of food to “laboring men” (as the census phrased it), setting B = 1 would imply that relief per day was sufficient to feed one adult male recipient per day. However, setting B = 1 is almost certainly too high, because adult males made up only a portion of recipients, and because “indoor relief (the almshouse) may have been cheaper to administer than outdoor relief.‘? In any case, the social statistics do not distinguish between outdoor and indoor relief, nor do they give information on the age and sex structure of relief
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recipients. If almshouse recipients accounted for 20% of total recipients (for whom 0 = 0.4; see Hannon, 1984b), while the remainder were mostly adults and their offspring (for whom we assume p = 0.9, on average), a plausible estimate of p is 0.8.” Setting p = 0.8, substituting into Equation 1 and re-arranging terms (because, for example, SLTN = 1 - SSTN - SSTF - SLTF), we estimate Equation 1 using weighted least squares (the weights are the county population totals; also note that the equation does not have an intercept term), pooling the data across the two years. The coefficients from the regression are the estimates of average duration in days per year, which we convert to months. The estimates appear in Table 3. On average, the duration of relief was much shorter among short-term recipients, as Lebergott (1976) defined them, particularly in the case of the foreign born. Although the paucity of other estimates makes comparisons hazardous, the durations seem consistent numerically with previous studies. Using a similar method, Hannon (198413) put the average duration of relief at 7 months in New York State in the 1820s. Ziliak’s (1995) recent study of relief in late nineteenth century Indianapolis pegs the average length of a complete spell at 8 months, but this may be a better estimate of duration among long-term recipients than among the general relief population. lg We use the estimated durations to construct a full-time equivalent measure of pauperism (see Hannon, 198413 for a similar use). The number of FTE paupers is pelt + 0.13 x pj, + 0.8 x p,, + 0.5 x pnsp where pq = the number of paupers in group i (f = foreign, n = native) of type j (st = short term, It = long term). The numbers multiplying the p,‘s are adjustment factors derived from the estimated durations; specifically, they are the ratios of the durations for type q relative to the duration for foreign-born long-term recipients. For example, the adjustment factor for short-term foreign recipients is 0.13 (= 1.5/l 1.5). Expenditures per FTE pauper is total expenditures on poor relief divided by the number of FTE paupers. Table 3.
Estimated Duration of Relief in Months Duration in Months
Short-term Long-term Short-term Long-term
1.5 11.5 4.6 8.5
Foreign Foreign Native Native
Source: See Appendix
Acknowledgment: participants Science
Comments
at the NBER,
Association
from
the Social
are gratefully
Rebecca
Blank,
Science
History
acknowledged.
Stanley
Engerman,
Association,
and conference
and the Allied
Social
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NOTES *Direct all correspondence to: Robert Margo, Vanderbilt University, Department of Economics, Nashville, TN 37235. 1. Gallman, personal correspondence. We divided Gallman’s GNP estimates (in constant 1860 dollars) for 1849 and 1859 by the federal census count of population in both years to derive the growth rate in the text. 2. The number of paupers was computed from DeBow (1854, p. 163) and Secretary of the Interior (1866, p. 512). All population figures are from U.S. Department of Commerce (1976). 3. Moreover, the majority of the additional recipients appear to have been ablebodied adults; see Hannon (1984b, p. 1015). 4. Specific instructions for collecting the wage data amounted to the statement that the “information called for in the six columns is so simple and so plainly set forth in the headings that it is deemed unnecessary to add thereto” (DeBow, 1854, pp. xxv). 5. Public records would not, however, have provided evidence on wages, also collected by the census; in this case, informed individuals probably provided estimates (see Margo, 1997, ch. 2). 6. That is, demographic shifts other than the percent foreign born, which both we (see section 3) and Hannon (198413) found were positively associated with pauperism. 7. Lebergott’s reasoning was that seasonal fluctuations in the demand for relief would generally have been at a low ebb in early June compared with, say, the winter months. Thus persons on relief in June were more likely to be long term relief recipientsthe mentally ill or blind, for example. As Hannon (1984b) points out, the majority of longterm recipients of relief were probably housed in poorhouses. 8. Using estimated durations as weights, Hannon (1984b, p. 1018) also shows that FTE pauperism increased at a much slower pace than total pauperism. 9. Similar models are discussed by Lebergott (1976) and Hannon (1996). 10. We assume in our econometric work that decisions are made at the county level although, strictly speaking, this is incorrect for Massachusetts, where decisions were made at the township level. 11. Specifically, suppose that e is set to maximize the utility of a decisive taxpayer, U(W - pe, e), where W is the wealth of the taxpayer, p is the pauper rate, and e is expenditures per pauper. Utility is maximized when PUl = lJ*. It is straightforward to show that@, < 0 and& > 0 provided that lJl2 is positive-that is, relief is a normal good. 12. By “full-time” we mean the daily wage of common labor, without board, multiplied by 311 days (see Lebergott, 1976). We did not include demographic variables other than percent foreign in X because, as already noted, previous work has found little evidence of demographic effects on pauper rates; see Hannon (1996). 13. There are 426 counties in the balanced sample. The sample underlying Table 1 is larger because it includes counties that were established between 1850 and 1860. 14. In her analysis of New York State administrative records, Hannon (1984a, p. 817) found that real expenditures per recipient fell between the late 1840s and late 1850s suggesting to her (1984b) that the “rise in relieved pauperism” did not occur because “the relief system became more generous”. However, as Panel B of Table 2 shows, a decline in expenditures could still be consistent with a greater willingness to support paupers-that is, an outward shift in the supply-of-relief function. The relative flatness of the demand for relief function implies that the shift in supply explains an
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economically significant fraction of the increase in FTE pauperism. To see this, note that the shift in supply accounts for 6S/(D + 68) of the increase in FTE pauperism, where D = demand shift, S = supply shift, and 6 is the ratio of the demand and supply coeffrcients of the expenditures variable (6 = 9.6). Using the figures reported in Table 2, supply shifts account for 62 percent [ = (9.6 x 0.75)/(4.4 + 9.6 x 0.75)] of the increase in the mean value of the FTE pauper rate. 15. For a comprehensive discussion of recent changes in poverty and welfare, see the various chapters in Danziger, Sandemr, and Weinberg (1994). 16. Unfortunately, the manuscripts for certain states that would be useful to study, such as New York, have been partially damaged (1850) or lost (for example, Ohio). 17. Adult males accounted for about 25 to 30% of total relief recipients, if data for New York State are any indication (see Hannon 198413, p. 1015). Assuming that the cost of board for adult women was the same as for adult men, the share of able-bodied adult recipients in New York was between 60-72% in the 1850s (Hannon, 198413, p. 1015). Adding in the elderly and disabled would raise the adult percentage to approximately 80%. 18. It should be noted that the estimates of duration are an implicit check on the reasonableness of the choice of 8. Specifically, the D’s cannot exceed 12 months (assuming the census data accurately measure a full year’s expenditures). Setting 8 to a lower number than 0.8 will produce longer durations; in particular, setting J3 = 0.6 produces an estimated duration exceeding 12 months for long-term foreign born recipients. 19. Ziliak argues that his data oversample seasonal relief cases, since he begins tracing individuals who entered the Indianapolis relief rolls during the first quarter of 1881. However, the proportion of short-term recipients among total recipients over the course of a year (the relevant period to compare with the census data) may be understated in Ziliak’s sample, since persons entering the rolls in, say, September, but leaving in October would be missed. Thus Ziliak’s estimate may better represent the mean duration of an (approximate) cross-section of relief recipients, which could be biased upwards relative to duration among all recipients.
REFERENCES Danziger, Sheldon H., Gary D. Sandefur, and Daniel H. Weinberg (Eds.). 1994. Confronting Poverty: Prescriptions for Change. Cambridge, MA: Harvard University Press. DeBow, J.D.B. 1854. The Seventh Census of the United States: 1850. Washington, DC: Robert Armstrong, Public Printer. Hannon, Joan. 1984a. “The Generosity of Antebellum Poor Relief,” Journal of Economic History, 44: 810-821. 1984b. “Poverty in the Antebellum Northeast: The View from New York State’s -. Poor Relief Rolls,” Journal of Economic History, 44: 1007-1032. 1996. Explaining Nineteenth Century Dependency Rates: Interplay of Life Cycles and -. Labor Markets. Unpublished paper, Department of Economics, St. Mary’s College. Katz. Michael. 1983. Poverty and Policy in American History. New York: Academic Press. Kiesling, Lynne. 1996. “Institutional Choice Matters: The Poor Law and Implicit Labor Contracts in Victorian Lancashire,“Ex@orations in Economic HistoT, 33: 65-85.
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Komlos, John. 1987. “The Height and Weight of West Point Cadets: Dietary Change in Antebellum America,” Journal of Economic History, 47: 897-927. Lebergott, Stanley. 1976. The American Economy: Income, Wealth, and Want. Princeton: Princeton University Press. Margo, Robert A. 1997. Antebellum Economic Performance: The Labor Market Perspective. Book manuscript in progress, Department of Economics, Vanderbilt University. Taylor, George Rogers. 1951. The Transportation Revolution. New York: Holt, Rinehart, and Winston, Inc. Trattner, Walter. 1994. From Poor Law to Welfare State: A History of Social Welfare in America. New York: Free Press. U.S. Secretary of the Interior. 1866. Statistics of the United States (including Mortality, Property, etc.) in 1860. Washington, DC: Government Printer. U.S. Department of Commerce. 1976. Historical Statistics of the United States. Washington, DC: USGPO. Wilentz, Sean. 1984. Chants Democratic. New York: Oxford University Press. Ziliak, Stephen. 1995. The Contradiction of Compassion: Private Charity and the “Solution” to Dependence. Unpublished paper, Department of Economics, University of Iowa.
Focus
Explaining the Rise in Antebellum Pauperism, 1850-1860: New Evidence
L. LYNNE KIESLING College of William & Mary
ROBERT
A. MARGO
Vanderbilt University
Between 1850 and 1860 the total ‘$auper rate”A”_
number of in&viduuls receiving public assistance per 1,000 population-increased j&n 5.8 to 10.2. We explore the determinants of the rise in antebellum pauperism using previously enexploited archival data. Changing labor market conditions, urbanization, and immigration led to a marked increase in the Lmund for public assistance. Antebellum taxpayers, however, were unwilling to maintain the generosity of relief at existing levels in the face of the rise in demand.
The 1850s were a decade of contrasting economic fortunes in the United States. According to Robert Gallman’s unpublished estimates, real per capita income increased by 24.5% between 1850 and 1860.’ Frontier states grew rapidly as population migrated westward in pursuit of economic gain. Railroad mileage increased by 360% over the decade, facilitating an enormous expansion in internal trade (U.S. Secretary of the Interior 1866, p. 331; Taylor, 1951). Against the backdrop of economic growth, the 1850s also witnessed a dramatic rise in the incidence of public “pauperism.” According to the 1850 census, the total pauper rate-the total number receiving poor relief from public funds during the year, per thousand persons- was 5.8/1,000 in 1850. B 1860 the total pauper rate had increased to 10.2/1,000, or by 76% in ten years. Y Although previous work has recognized the increase in pauperism during the 185Os, there is no consensus on the relative economic significance of different factors (Hannon, 198413, 1996; Trattner, 1994). This paper uses previously untapped archival evidence, in conjunction with published census data, to explore the determinants of public pauperism during the 1850s. The data, which are derived from the manuscript censuses of social statistics of 1850 and 1860, are sufficiently rich to permit us to estimate an econometric model of pau-
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per rates at the county level. Our principal findings are that changing labor market conditions, urbanization, and immigration led to a marked increase in the demand for public assistance between 1850 and 1860. However, antebellum taxpayers were apparently unwilling to maintain the existing level of financial generosity of public assistance in the face of increased demand.
I.
BACKGROUND
DISCUSSION
Since the New Deal the provision of public welfare in the United States has been the primary responsibility of the federal government. Recent reforms, however, are in the process of shifting that responsibility back to state and local governments. In decentralizing welfare policy, current reform efforts are gradually returning the system to its historical roots-that is, to a system in which “local”, as opposed to “national”, governments paid for, and implemented, relief policy. Antebellum poor relief was a diverse affair. This diversity reflects the local nature of the English poor law, from which American poor relief policy was derived during the colonial period (Trattner, 1994). American colonial statutes preserved the family responsibility for the needy, with public relief available for those unable to care for themselves, or whose families were unable to do so. During the antebellum period local governments (primarily at the county level) were fiscally responsible for providing relief. Recipients generally had to meet some form of residency requirement. Public provision of relief included the poorhouse (indoor relief), and cash or in-kind transfers (outdoor relief). Public officials favored poorhouses, on the belief that outdoor relief had more severe disincentive effects on work effort (Trattner, 1994). Contemporary observers recognized that pauperism was increasing in the 1850s. Although many attributed rising pauperism to moral failings on the part of the poor, the traditional view among historians places the blame for the rise in pauperism squarely on urbanization, industrialization, and immigration (Hannon, 1984b, p. 1018). Urbanization was at fault because the urban poor allegedly lacked access to the social networks (family and friends) that traditionally sustained the rural poor through difficult times. The displacement of the artisanal shop with the factory system and with “outwork” reduced the skill requirements of industrial labor and with it, household income, putting industrial workers at greater risk of needing occasional public assistance (Wilentz, 1984). Immigration was a culprit for reasons similar to industrialization-the foreign born had lower incomes than the native born and, therefore, more prone to seek relief when faced with an adverse economic shock. Recent research has provided a more nuanced view of the possible causes of rising antebellum pauperism (Katz, 1983; Hannon, 198413, 1996). Labor demand, rural and urban, was highly seasonal during the antebellum period. In rural areas, however, under-employment was more readily disguised, and farm households could generally meet their consumption needs through home pro-
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duction. But, with the spread of wage labor (via industrialization and urbanization), workers became more dependent on market-generated incomes. Unable to engage in home production during bouts of seasonal (or cyclical) unemployment, the urban poor increasingly turned to public assistance. Lacking the savings required to smooth consumption, low-wage workers were especially vulnerable when, as in the 185Os, the price of food relative to unskilled wages rose sharply (Komlos, 1987, p. 915; Margo, 1997, ch. 2). Analysis of administrative data for New York State by Joan Hannon (1984b) provides some support for this nuanced view of the antebellum welfare explosion. Between the late 1840s and late 1850s pauper rates in New York state rose by nearly 60%. Yet residency rates in poorhouses remained roughly constant, implying that the vast majority of the additional paupers were receiving outdoor relief for relatively brief periods of time (Hannon, 198413, pp. 1012, 1018).3 Although Hannon’s data did not permit her to examine directly the relationship between low wages and pauperism, she was able to show that county-level pauper rates were negatively correlated with the extent of household production, and positively with urbanization and the percent foreign born (Hannon, 1984b, p. 1026).
II.
ARCHIVAL
DATA ON POOR RELIEF
Although various states produced statistics on the incidence and cost of public poor relief beginning in the early 18OOs, the first attempt to produce comparable national statistics on poor relief occurred in 1850. As part of the enumeration effort in 1850 and again in 1860, census marshals were instructed to collect information for minor civil divisions (MCDs) on the total number of recipients of poor relief (native and foreign); the number receiving poor relief on June 1 (native and foreign); and total expenditures on poor relief. At the time the census was compiled, the data were aggregated to the state level, and state totals were reported in the published volumes of the 1850 and 1860 censuses. Economic historians have used these state totals in previous work on antebellum poor relief (Lebergott, 1976). The original manuscripts of the social statistics survive for some states, however, and microfilm copies are available at the National Archives or from various state archives. For the purposes of this paper we obtained and computerized data from microfilms for eight states, two from each census region: Pennsylvania and Massachusetts (Northeast); Michigan and Iowa (Midwest); North Carolina and Virginia (South Atlantic); and Kentucky and Tennessee (South Central). The selection of states was dictated by the cost of data retrieval, balanced against the desire to have a regionally dispersed sample of counties. The instructions to census enumerators specified that social statistics were to be collected for each MCD (or townships, in the case of New England) “as far as practicable” (DeBow, 1854, pp. xxiv). Inspection of the microfilms suggest that
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this goal was broadly achieved, in the sense that virtually all counties reported data for at least one MCD. Sometimes the reported MCD names correspond to known geographic entities (for example, wards in Philadelphia) but in other cases they do not (for example, a census marshal might refer to an MCD as “my division”). Because of this problem (along with illegibility of some MCD names) we have chosen to aggregate all data to the county level. Other than commenting that information was “not to be ascertained entirely by personal inquiry of individuals, but in part from public records and reports, and public offices” the census was silent on the exact procedures marshals were to follow in collecting the information on poor relief (Debow, 1854, p. xxiv).4 Unfortunately, none of the microfilms contain marginalia on data collection procedures. However, “public records” probably provided most or all of the relevant figures, on the assumption that individuals chosen randomly by the census enumerators would presumably be poorly informed about the number of persons on relief or the amount spent on relief.5 The problems with the data are numerous. No information is available on the type of relief (outdoor versus indoor), or on age or gender characteristics, although Hannon’s (198413) analysis of New York administrative records suggests that the increase in pauperism in the 1850s was not caused by demographic shifts.6 Direct information on the average duration of relief (for example, in days or months) was not reported. Lebergott (1976), however, suggested that the number of recipients on June 1 was a good proxy for the “long-term” relief population.’ The implication of Lebergott’s suggestion is that the remaining recipients (Total recipients - June 1 recipients) can be thought of as “shortterm.” The appendix to the paper develops a method to infer the average duration of relief from the census distinction between short and long-term recipients, and our estimates of average duration are consistent with Lebergott’s interpretation-the predicted duration of relief was considerably shorter for “short-term” recipients (as Lebergott defined them) than “long-term” recipients (see Appendix Table 3). This study uses our estimates of duration to construct “lull-time equivalent” (FTE) measures of pauperism. In addition, many MCD’s, particularly in frontier areas (for example, various counties in Iowa in 1850) simply left the relevant columns blank, which we have interpreted as true zeros. Upon aggregating to the county level, nonreporting is reduced but not eliminated. However, once the data are weighted by countylevel population counts, the economic significance of “zeros” is greatly reduced. All the numerical estimates reported in this paper are weighted by relevant county-level populations. Sample statistics on pauper rates are reported in Table 1. Consistent with the direction of change revealed by the published census aggregates, the total pauper rate in the sample increased between 1850 and 1860 from 4.6/1000 to 7.8/1000, although the levels of pauperism in the sample are below (in both years) the aggregate levels reported in the published census. Consistent with Hannon’s (1984b) findings for New York State, the rise in the total pauper rate
EXPLAINING
Table1. Short-term Long-term Total FI’E Total
THE BISE IN ANTEBELLUM
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Sample Statistics: Paupers per 1,000 Population 1850
1860
2.5 2.1 4.6 2.2
5.4 2.4 7.9 3.2
Soum: Eight-state sample fmm manuscript censuses of social statistics.
between 1850 and 1860 was dominated by the rise in short-term relief. Because the duration of relief among short-term recipients was shorter than among longrun recipients, the increase in ETE pauperism between 1850 and 1860 was less than the increase in the total pauper rate. The final row of Table 1 computes FTE pauper rates for the sample, using the estimated durations reported in the appendix as weights (see the appendix). In FTE terms, the antebellum welfare explosion was less dramatic-instead of a 70% (= 7.8/4.5 - 1) increase, the FTE rate rose by 45% (= 3.2/2.2 - I).’
III.
A MODEL OF ANTEBELLUM
POOR REFIEF
To motivate our empirical analysis this section sketches a simple model of public poor relief. The model consists of two functions: a “demand-for-relief’ equation and a “supply-of-relief’ equation.g Both demand and supply are determined at the local level, which for the purposes of our analysis is taken to be the county.” The demand for public poor relief is p = d(X, e), where p is the incidence of public pauperism (for example, the number of paupers per 1,000 population); X is a set of exogenous variables that shift the demand for relief; and e is the “generosity” of relief, assumed to be measured by expenditures per recipient. Following the previous discussion, both p and e are expressed in FTE units. By “demand” we mean that pauperism and generosity are positively related-holding other factors (X) constant, the more generous is relief, the greater the number of persons seeking relief (d, > 0). The supply-of-relief equation is p = ~(2, e), where 2 is a set of variables that shift the supply function. We think of supply as the outcome of an maximization process in which antebellum taxpayers agree to levy local taxes (for example, on property) to fund a socially optimal level of public relief. The exact motives for supplying relief are left unspecified, although at some level we presume that taxpayers are altruistic towards the poor- that is, public assistance was a normal good. Because the marginal utility of helping those is need is assumed to be decreasing, however, and because taxpayer resources are finite, there is a negative tradeoff between p and e on the supply side-holding constant 2, taxpayers are less willing to be generous (se < 0) as p increases. * ’ Equilibrium is reached when p and e are consistent with each other-that is, at a given level of e, the number of paupers desiring relief just equals the num-
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ber that taxpayers are willing to support. Note that if various factors change to cause the demand for relief to increase, p will rise but e will fall. Conversely, if taxpayers become more willing to support (for example, as their wealth increases) p will also rise but so will e. For estimation purposes we assume that bothf and g are linear: Demand: p = xa
+ J3e + E/j
Supply: p = zr - Se + E, where a, J3. I-, and 6 are parameters, E’S are random errors. Included in X are four variables: the “real unskilled wage,” defined to be the full-time annual wage of unskilled labor without board, deflated by the cost of board (food); a dummy variable indicating the presence of an urban area in the county of population 10,000 or more; the percent foreign born; and the number of churches per capita.‘* Like Hannon (1984b) we do not have a direct measure of the extent of wage labor. However, if low-wage workers were especially vulnerable to economic distress, as Hannon suggests, we should observe a strong negative relationship between the real unskilled wage and the FTE pauper rate. For the reasons discussed in Section II, the coefficients of the urban dummy and the percent foreign are expected to be positive. Recent work by Kiesling (1996) and Ziliak (1995) suggests that private and public sources of relief were substitutes. The number of churches per capita is included as a proxy for the availability of private relief; its coefficient is expected to be negative. Included in 2 are the urban dummy, churches per capita, and taxable wealth per capita. The urban dummy is included in the supply equation because the cost of providing relief may have differed between urban and rural counties. If, for example, the urban cost was lower, the coefficient of the urban dummy should be positive. Just as the availability of private sources may have lowered the likelihood of those in need to seek public assistance, the willingness to supply public funds may have been reduced. However, a large number of churches per capita may also capture a heightened sense of social responsibility towards the poor-that is, a shift in community tastes for providing more generous relief. Thus, the coefficient of churches per capita in the supply equation is uncertain. The principal variable shifting the supply equation is taxable wealth per capita. If public relief was a normal good, then higher taxable wealth should have increased the supply of relief. Thus the coefficient of taxable wealth should be positive. To maintain consistency with the definition of the real wage variable, wealth per capita and expenditures per recipient are deflated by the cost of board. The wage of unskilled labor, the cost of board, and taxable wealth are drawn from the manuscript censuses of social statistics, which reported these data along
EXPLAINING
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with the those on poor relief. All other data are drawn from the published volumes of the 1850 and 1860 censuses. To complete the specification we include county fixed effects in both the supply and demand equations. We restrict the estimation to a balanced sample of counties-that is, counties that existed in both 1850 and 1860 and which reported complete data. l3 Because we have excluded wealth per capita from the demand equation, and the real wage and the percent foreign from the supply equation, both equations are identified. Thus, after first-differencing, we estimate the equations using weighted two-stage least squares. The weights are the average of the population counts in 1850 and 1860. Sample means and coefficients are presented in Panel A of Table 2. Both regressions, as well as the majority of the coefficients are statistically significant, and the signs of coefficients accord with our prior expectations. On the demand side, labor market conditions evidently mattered, as indicated by the negative coefficient on the real unskilled wage. So, too, did urbanization and immigration: counties that underwent urban expansion during the decade, as well as immigration, witnessed an expansion in the demand for poor relief. Counties with a greater density of churches appear to have had lower demand for public relief, consistent with the belief that public and private assistance were substitutes. On the supply side, there is little evidence that the cost of supplying poor relief was lower in urbanized counties, or that the presence of a large number of churches per capita significantly altered the supply of public assistance. Richer counties, however, clearly were more willing and able to provide public relief, as evidenced by the positive and significant coefficient of per capita wealth in the supply equation. We also find strong evidence, on both the demand and supply side, of a significant relationship between the incidence of pauperism and of the generosity of relief. As hypothesized in our model, pauper rates were a positive function of expenditures per recipient on the demand side, and a negative function of expenditures on the supply side. The supply equation was much steeper, in relative terms, than the demand equation-the number of paupers seeking relief was quite responsive to the generosity of relief, while local governments imposed a sharp negative tradeoff between pauperism and generosity. In Panel B of Table 2 we report the results of a decomposition analysis, in which we use the sample mean values of the exogenous variables to compute the mean values of the supply and demand shift terms. The mean value of the demand shift term was 4.4, which implies that the demand for relief rose, in FTE terms, by 4.4 per 1,000 between 1850 and 1860. However, the actual increase in the FTE pauper rate in the sample counties was 1.67. Although labor market conditions, urbanization, and immigration served to substantially raise the demand for relief in the 185Os, the steepness of the negative tradeoff between pauperism and generosity on the supply side limited the overall increase in pauperism. l4 Antebellum taxpayers were apparently unwilling to
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Demand-Supply Model of Antebellum Pauperism Sample Mean
A. Weighted TSLS Fixed Effects First Difference of:
Estimates,
Real Unskilled Wage
-0.065
Percent Foreign
0.017
Churches/Population
0.083
Urban County
0.068
Real Wealth/Population
0.083
per FTE Pauper
F-Statistic Mean value of dependent Number of Counties
SUPPlY
1.581 (6.040) -20.389) (4.53 1) 52.158 (4.025) -1.943 (3.321) 12.114 (4.227)
0.745 (3.079)
First Differences
Constant
Real Expenditures
Demand
variable
-0.103
(4.552) -3.390 (2.582) 5.69
1.09 426 Demand
B. Decomposition of Demand and Supply Shifts Predicted Shift, 1850-60 Due to: Real Wage Percent Foreign Urbanization Churches per Capita Wealth per Capita
32.549 (4.349) 4.58
-0.320 (0.678) -1.446 (1.397) 1.554
4.44 1.35 0.89 0.82 -0.20
SUPPlY 0.75
-0.10 -0.03 0.13
Notes: 1.
Dependent variable is first difference (1860-1850) of FTE pauper rate (number of FTE paupers per 1,000 population). Pauper rate and expenditures per pauper are endogenous variables. Absolute value oftstatistics in parentheses. Weight is average of 1850 and 1860 population count. Source: eight-state sample from manuscript censuses of social statistics. 2. Predicted Shift equals sum of mean values of exogenous variables times regression coefficients (from Panel A); see text.
maintain the generosity of poor relief at 1850s levels, on average, when faced with the sharply increased demand for public assistance between 1850 and 1860. IV.
CONCLUDING
REMARKS
Despite rising per capita income, the fraction of persons receiving public assistance in the United States rose substantially between 1970 and 1990. Although some scholars attribute the post-1970 rise in welfare incidence to overly generous public assistance, others point to labor market factors, such as immigration and to decreased relative demand for less-educated workers.t5 These modern
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concerns echo those of the lS5Os, which witnessed one of the first. welfare explosions in American history. Labor market conditions, immigration, and structural change led to a marked increase in the demand for public assistance between 1850 and 1860. Antebellum taxpayers were apparently willing to help those in genuine need, but their generosity was strained by the rise in demand and, cons~q~endy, the supply of relief did not keep pace. The results of the paper could be extended in two directions. First, it would be valuable to expand the sample to include additional statesi Second, it would be useful to estimate a similar econometric model for a later (but pre-New Deal) period (if suitable data coulcl be found) to see if the magnitudes of the tradeoffs between pauperism and expenditures on the supply and demand side remained stable over time.
APPENDIX Estimating the Average
This duration
appendix of relief.
Duration of Relief
develops a method for using the census data to estimate Our estimates are derived from the following equation:
the
where
ed= Dg i j Sq i j
= = = = = =
expenditures per recipient per day; average duration of relief, in days, L1‘ (long-term), ST (short term), ~(ative), ~(forei~~); share of relief type in total recipients, LT, ST, N, F.
The data contain no direct information on cd. In place of direct infor~~ation, we make the assumption that ed = ~*(we~kly cost of board/i’). 7hat is, expenditures per day were a constant fraction of the daily cost of food. Because the data on board refer to the cost of food to “laboring men” (as the census phrased it), setting B = 1 would imply that relief per day was sufficient to feed one adult male recipient per day. However, setting B = 1 is almost certainly too high, because adult males made up only a portion of recipients, and because “indoor relief (the almshouse) may have been cheaper to administer than outdoor relief.‘? In any case, the social statistics do not distinguish between outdoor and indoor relief, nor do they give information on the age and sex structure of relief
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recipients. If almshouse recipients accounted for 20% of total recipients (for whom 0 = 0.4; see Hannon, 1984b), while the remainder were mostly adults and their offspring (for whom we assume p = 0.9, on average), a plausible estimate of p is 0.8.” Setting p = 0.8, substituting into Equation 1 and re-arranging terms (because, for example, SLTN = 1 - SSTN - SSTF - SLTF), we estimate Equation 1 using weighted least squares (the weights are the county population totals; also note that the equation does not have an intercept term), pooling the data across the two years. The coefficients from the regression are the estimates of average duration in days per year, which we convert to months. The estimates appear in Table 3. On average, the duration of relief was much shorter among short-term recipients, as Lebergott (1976) defined them, particularly in the case of the foreign born. Although the paucity of other estimates makes comparisons hazardous, the durations seem consistent numerically with previous studies. Using a similar method, Hannon (198413) put the average duration of relief at 7 months in New York State in the 1820s. Ziliak’s (1995) recent study of relief in late nineteenth century Indianapolis pegs the average length of a complete spell at 8 months, but this may be a better estimate of duration among long-term recipients than among the general relief population. lg We use the estimated durations to construct a full-time equivalent measure of pauperism (see Hannon, 198413 for a similar use). The number of FTE paupers is pelt + 0.13 x pj, + 0.8 x p,, + 0.5 x pnsp where pq = the number of paupers in group i (f = foreign, n = native) of type j (st = short term, It = long term). The numbers multiplying the p,‘s are adjustment factors derived from the estimated durations; specifically, they are the ratios of the durations for type q relative to the duration for foreign-born long-term recipients. For example, the adjustment factor for short-term foreign recipients is 0.13 (= 1.5/l 1.5). Expenditures per FTE pauper is total expenditures on poor relief divided by the number of FTE paupers. Table 3.
Estimated Duration of Relief in Months Duration in Months
Short-term Long-term Short-term Long-term
1.5 11.5 4.6 8.5
Foreign Foreign Native Native
Source: See Appendix
Acknowledgment: participants Science
Comments
at the NBER,
Association
from
the Social
are gratefully
Rebecca
Blank,
Science
History
acknowledged.
Stanley
Engerman,
Association,
and conference
and the Allied
Social
EXPLAINING THE RISE IN ANTEBELLUM PAUPERISM, 1850-1860
415
NOTES *Direct all correspondence to: Robert Margo, Vanderbilt University, Department of Economics, Nashville, TN 37235. 1. Gallman, personal correspondence. We divided Gallman’s GNP estimates (in constant 1860 dollars) for 1849 and 1859 by the federal census count of population in both years to derive the growth rate in the text. 2. The number of paupers was computed from DeBow (1854, p. 163) and Secretary of the Interior (1866, p. 512). All population figures are from U.S. Department of Commerce (1976). 3. Moreover, the majority of the additional recipients appear to have been ablebodied adults; see Hannon (1984b, p. 1015). 4. Specific instructions for collecting the wage data amounted to the statement that the “information called for in the six columns is so simple and so plainly set forth in the headings that it is deemed unnecessary to add thereto” (DeBow, 1854, pp. xxv). 5. Public records would not, however, have provided evidence on wages, also collected by the census; in this case, informed individuals probably provided estimates (see Margo, 1997, ch. 2). 6. That is, demographic shifts other than the percent foreign born, which both we (see section 3) and Hannon (198413) found were positively associated with pauperism. 7. Lebergott’s reasoning was that seasonal fluctuations in the demand for relief would generally have been at a low ebb in early June compared with, say, the winter months. Thus persons on relief in June were more likely to be long term relief recipientsthe mentally ill or blind, for example. As Hannon (1984b) points out, the majority of longterm recipients of relief were probably housed in poorhouses. 8. Using estimated durations as weights, Hannon (1984b, p. 1018) also shows that FTE pauperism increased at a much slower pace than total pauperism. 9. Similar models are discussed by Lebergott (1976) and Hannon (1996). 10. We assume in our econometric work that decisions are made at the county level although, strictly speaking, this is incorrect for Massachusetts, where decisions were made at the township level. 11. Specifically, suppose that e is set to maximize the utility of a decisive taxpayer, U(W - pe, e), where W is the wealth of the taxpayer, p is the pauper rate, and e is expenditures per pauper. Utility is maximized when PUl = lJ*. It is straightforward to show that@, < 0 and& > 0 provided that lJl2 is positive-that is, relief is a normal good. 12. By “full-time” we mean the daily wage of common labor, without board, multiplied by 311 days (see Lebergott, 1976). We did not include demographic variables other than percent foreign in X because, as already noted, previous work has found little evidence of demographic effects on pauper rates; see Hannon (1996). 13. There are 426 counties in the balanced sample. The sample underlying Table 1 is larger because it includes counties that were established between 1850 and 1860. 14. In her analysis of New York State administrative records, Hannon (1984a, p. 817) found that real expenditures per recipient fell between the late 1840s and late 1850s suggesting to her (1984b) that the “rise in relieved pauperism” did not occur because “the relief system became more generous”. However, as Panel B of Table 2 shows, a decline in expenditures could still be consistent with a greater willingness to support paupers-that is, an outward shift in the supply-of-relief function. The relative flatness of the demand for relief function implies that the shift in supply explains an
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economically significant fraction of the increase in FTE pauperism. To see this, note that the shift in supply accounts for 6S/(D + 68) of the increase in FTE pauperism, where D = demand shift, S = supply shift, and 6 is the ratio of the demand and supply coeffrcients of the expenditures variable (6 = 9.6). Using the figures reported in Table 2, supply shifts account for 62 percent [ = (9.6 x 0.75)/(4.4 + 9.6 x 0.75)] of the increase in the mean value of the FTE pauper rate. 15. For a comprehensive discussion of recent changes in poverty and welfare, see the various chapters in Danziger, Sandemr, and Weinberg (1994). 16. Unfortunately, the manuscripts for certain states that would be useful to study, such as New York, have been partially damaged (1850) or lost (for example, Ohio). 17. Adult males accounted for about 25 to 30% of total relief recipients, if data for New York State are any indication (see Hannon 198413, p. 1015). Assuming that the cost of board for adult women was the same as for adult men, the share of able-bodied adult recipients in New York was between 60-72% in the 1850s (Hannon, 198413, p. 1015). Adding in the elderly and disabled would raise the adult percentage to approximately 80%. 18. It should be noted that the estimates of duration are an implicit check on the reasonableness of the choice of 8. Specifically, the D’s cannot exceed 12 months (assuming the census data accurately measure a full year’s expenditures). Setting 8 to a lower number than 0.8 will produce longer durations; in particular, setting J3 = 0.6 produces an estimated duration exceeding 12 months for long-term foreign born recipients. 19. Ziliak argues that his data oversample seasonal relief cases, since he begins tracing individuals who entered the Indianapolis relief rolls during the first quarter of 1881. However, the proportion of short-term recipients among total recipients over the course of a year (the relevant period to compare with the census data) may be understated in Ziliak’s sample, since persons entering the rolls in, say, September, but leaving in October would be missed. Thus Ziliak’s estimate may better represent the mean duration of an (approximate) cross-section of relief recipients, which could be biased upwards relative to duration among all recipients.
REFERENCES Danziger, Sheldon H., Gary D. Sandefur, and Daniel H. Weinberg (Eds.). 1994. Confronting Poverty: Prescriptions for Change. Cambridge, MA: Harvard University Press. DeBow, J.D.B. 1854. The Seventh Census of the United States: 1850. Washington, DC: Robert Armstrong, Public Printer. Hannon, Joan. 1984a. “The Generosity of Antebellum Poor Relief,” Journal of Economic History, 44: 810-821. 1984b. “Poverty in the Antebellum Northeast: The View from New York State’s -. Poor Relief Rolls,” Journal of Economic History, 44: 1007-1032. 1996. Explaining Nineteenth Century Dependency Rates: Interplay of Life Cycles and -. Labor Markets. Unpublished paper, Department of Economics, St. Mary’s College. Katz. Michael. 1983. Poverty and Policy in American History. New York: Academic Press. Kiesling, Lynne. 1996. “Institutional Choice Matters: The Poor Law and Implicit Labor Contracts in Victorian Lancashire,“Ex@orations in Economic HistoT, 33: 65-85.
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