Hostname: page-component-78c5997874-t5tsf Total loading time: 0 Render date: 2024-11-17T10:21:40.520Z Has data issue: false hasContentIssue false
  • English
  • Français

The Distribution of the Income in the Great Depression: Preliminary State Estimates

Published online by Cambridge University Press:  03 March 2009

Mark Schmitz  and
Price V. Fishback
Mark Schmitz
Affiliation:
Associate Professor of Finance and Business Economics, University of Washington, Seattle, Washington 98195
Price V. Fishback
Affiliation:
Associate Professor of Finance and Business Economics, University of Washington, Seattle, Washington 98195
Get access Rights & Permissions [Opens in a new window]

Abstract

State-level estimates of income shares for the top one and five percent of the population are presented for 1929, 1933, and 1939. Significant cross-sectional variation is found in 1929, but the range narrows as the shares fall dramatically to 1933. Analysis indicates that property incomes influence the shares but provides little evidence of a tradeoff between per capita income and inequality as measured by the shares.

Type
Papers Presented at the Forty-Second Annual Meeting of the Economic History Association
Information
The Journal of Economic History , Volume 43 , Issue 1 , March 1983 , pp. 217 - 230
Copyright
Copyright © The Economic History Association 1983

Access options

Get access to the full version of this content by using one of the access options below. (Log in options will check for institutional or personal access. Content may require purchase if you do not have access.)

References

The authors appreciate the helpful comment of Robert Higgs and Lee Alston. Computational assistance was provided by M. Schmitz and E.Schmitz.Google Scholar

1 For an overview of state per capita income, see Creamer, Daniel and Merwin, Charles, “State Distribution of Income Payments, 1929–1946,” Survey of Current Business (July 1942), pp. 18–26;Google Scholar and Hanna, Frank, State Income Differentials, 1919–1954 (Durham, North Carolina, 1959).Google Scholar

2 Kuznets, Simon, Shares of Upper Income Groups in Income and Savings (New York,1954).Google Scholar

3 Mendershausen, Horst, Changes in Income Distribution During the Great Depression (New York, 1946), pp. 114–15.Google ScholarA summary of the evidence can be found in Williamson, and Lindert, , American Inequality: A Macroeconomic History (New York, 1981), pp. 75–80, 315–18,Google ScholarThe distribution of income within the top groups is also examined by Tucker, Rufus, “The Distribution of Income Among Income Taxpayers in the United States, 1863–1935”, Quartely Journal of Economics, 52 (08 1938), 547–87.CrossRefGoogle ScholarSee also, Goldsmith, Selma et al. , “Size Distributions of Income Since the Mid-Thirties,” Review of Economics and Statistics, 36 (02 1954), 1–32.CrossRefGoogle Scholar

4 For a provocative study using this approach with the Gini coefficient see Fei, John, Ranis, Gustav, and Kuo, Shirley, Growth with Equity: The Taiwan Case (New York, 1979). Chapters 8–12 provide the mathematical exposition.Google Scholar

5 Kuznets, Shares, pp. 63–92.Google ScholarMoore, Geoffrey H., “Secular Changes in the Distribution of Income,” American Economic Review, 42 (05 1952), 524–44.Google Scholar

6 We follow the Department of Commerce breakdown of income payments. Briefly, “wage and salaries” is labor payments net of retirement and unemployment contributions. Proprietor income includes “incomes of self-employed persons available for personal use” or “entrepreneurial withdrawals.” Farm sources are included but imputed rents are not in this flow nor the total. Dividends, interest, rents, and royalties comprise property incomes and are based partially upon income tax returns. A final category is ‘other labor income’ or simply ‘other income,’ which includes pension and relief payments.Google Scholar

7 See Kuznets, Simon, “Economic Growth and Income Inequality’, American Economic Review, 45 (03 1955), 1–28;Google Scholarand “Quantitative Aspects of the Economic Growth of Nations: VIII, Distribution of Income by Size,” Economic Development and Cultural Change, 11 (Jan. 1963), 1–80.Google ScholarIn the 1920s and 1930s a large portion of high incomes were reported in urban areas. For example, in 1934 (hardly a good year, unfortunately) slightly under half of the returns with over $25,000 net income were found in the states of New York, Pennsylvania, and Illinois. Over 40 percent of those returns were reported in the ten largest cities. It is possible of course, that some of the reported urban incomes were from rural property holdings or payments generated in other locations. Returns by location can be found in U. S. Department of Treasury, Division of Research and Statistics, Statistics of Income Supplement, 1934 (Washington, D.C., 1938), but are not available for earlier years.Google Scholar

8 While property incomes declined drastically during the Depression the flow's share in total income actually rose in many states from 1929 to 1933 (see Table 3 below). This is partly the result of the exclusion of capital gains from property incomes. Net capital gains listed on 1929 tax returns would have added 25 percent to property income and 5 percent to total income. The 90 percent decline in net gains to 1933 is much greater than the fall in other income sources.Google Scholar

9 Mendershausen, Changes, pp. 73–77, 116.Google Scholar

10 Williamson and Lindert, American Inequality, pp. 285–90.Google ScholarCreamer, Daniel, Personal Incomes During Business Cycles (Princeton, 1956), pp. 34–59, 113–14.Google Scholar

11 The income-equality tradeoff is summarized in Kuznets, “Quantitative Aspects.” See also Adelman, Irma and Morris, Cynthia Taft, Economic Growth and Social Equity in Developing Countrie (Stanford, 1973).Google ScholarNumerous studies of individual nations support the hypothesis across regions and time with the notable excption being the Taiwan case discussed in Fei, et al., Growth with Equity, pp. 312–323.Google Scholar

12 Additional variables have been used to examine interstate variations in the distribution of income after survey and census data on incomes became available. For examples, see Al-Samarrie, Ahmad and Miller, H. P., “State Differentials in Income Concentration,” American Economic Review, 57 (03 1967) 59–72;Google ScholarD. J. Aigner and A. J. Heins, “On the Determinants of Income Inequality,” Ibid., pp. 175–184; and Verway, David, “A Ranking of States by Inequality Using Census and Tax Data,” Review of Economics and Statistics, 48 (08 1966), 314–21.CrossRefGoogle ScholarKuznets, “Quantitative Aspects,” pp. 36–44, 74–79 also examines state distributions estimated by Seymour Goodman for the 1950s.Google Scholar

13 U.S. Treasury Department, Bureau of Internal Revenue, Statistics of Income. Annual report.Google Scholar

14 The earliest estimates are reported in Nathan, Robert R. and Martin, John L., State Income Payments, 1929–1937 (Washington, D.C., 05 1939).Google ScholarSubsequent (and periodically revised) estimates were regularly reported in the Survey of Current Business; see April 1940, October 1940, August 1941, July 1942, July 1943, and the August issues for 1944–1954. The payments series was replaced by Personal Income in 1955. Personal income is also available to 1929 but conforms less closely to the gross income estimates from tax returns.Google Scholar

15 Kuznets, Shares, p. 35. For a number of states with higher income levels (e.g., New York, Illinois, and Massachusetts) it is possible to estimate shares further into the income distribution.Google Scholar

16 Kuznets, Shares, Table III, pp. 516–18. The number of persons per return (PPR) can be calculated across states or income classes but not jointly across both. Variation in PPR by class is small except for lower income groups ($3,000 and under); state PPRs range from 1.8 to 2.3. State variations result from differences in the distribution of returns by income class, the number of dependents per joint return, and the ratio of joint to single returns. We assumed that the first factor is the crucial one and accordingly used the national PPRs by income class for all states rather than the state PPR for every income class in the state. We examined the potential bias from this assumption by comparing the estimated population covered by all returns using this method with the calculated return population for the states. For states with high PPRs (e.g., southern states) our method underestimated the population, consequently underestiating the number of returns needed to find one percent of the population as well as the income for the group. The opposite holds for the below average PPR states. The maximum potential error appears to be about 0.5 percent for ONE. Only a few states, however, are affected to this degree.Google Scholar

17 The desired denominator is gross income payments to residents by state rather than by location of the payer. The difference between the two concepts is probably greatest for wages and salaries. Estimates were based on establishment data, and wages of non-residents would therefore be incorrectly allocated. The Commerce Department adjusted totals for New York, New Jersey, Maryland, and Virginia to reduce this discrepancy. Proprietor and property (capital) payments are based on individual censuses (including the tax returns) and are affected to a lesser degree. A different conceptual problem is that the functional distribution of income by state is mismeasured when property income generated in a state is not listed in that state's production because it is received by a non-resident. For comments on these problems and the wage and salary adjustments see, Creamer and Merwin, ‘State Income Payments,’ pp. 19, 26.Google Scholar

18 George Stigler provides similar estimates for 1940 using a different gross to net transformation and a broader definition of income. See his Trends in Employment in the Service Industries (Princeton, 1956), pp. 96–97.Google Scholar

19 Creamer and Merwin, “State Income Payments,” pp. 23–26.Google Scholar

20 Delaware and North Dakota are omitted from our analysis. For Delaware, 29 percent of 1929 net income was in only two returns and the Commerce Department frequently mentions difficulties in estimating and using the state's income statistics. North Dakota has been excluded because of errors in the 1929 published tax data, pointed out in Kuznets, Shares, p. 249.Google Scholar

21 We should caution the reader that ONE and FIVE can only provide a partial view of the distribution of income within a state. Broader measures available after World War II (see the references in footnote 12 above) rank some western and southern states relatively high in inequality.Google Scholar

22 Louis Cain has raised the interesting point that the large tax increase of 1932 might affect our estimates if it caused high income individuals to shift from taxable to nontaxable income sources. This would reduce reported taxable income and reduce ONE even if the group's total income was unchanged. Gene Smiley discusses a possible reverse effect from the 1924 and 1925 tax cuts in his “Did Incomes for Most of the Population Fall from 1923 through 1929?” included in this volume. We partially examined this hypothesis by making annual estimates for four states where we had found large declines in ONE. In each case the drop occurred quite rapidly and before the tax increase:Google Scholar

23 Mendershausen, Changes, pp. 54–56. See also Creamer, Personal Income, pp. 60–76.Google Scholar

24 The relationship should actually be between ONE and real PERCAP but we have not attempted a hazardous and unappealing adjustment for price level differences. Nonetheless it is likely that nominal differences are greater than real ones because nominal PERCAP is certainly positively correlated with state price levels. Our reported coefficients are therefore upwardly biased estimates of the true coefficients for real PERCAP. Further examination of the ONE and PERCAP estimates also indicates no evidence for all or a portion of a U-shaped relationship between the two.Google Scholar

25 We can also interpret the coefficients as estimates of the share of each flow that went to the top 1 percent. Because the regressions are not weighted, however, the estimated coefficients are biased.Google Scholar

26 Simila r results were found with equations using nonwage (proprietor and property) income rather than property alone. One of our concerns was that property incomes were reported in states other than where the income was generated, biasing the property share estimates. We felt this to be particularly ominous for states such as New York, New Jersey, and Massachusetts. To test for the effects of a reporting bias on our slope coefficients we defined a dummy variable equal to one for those states and zero otherwise, and reran each equation with the dummies entered interactively with the independent variables. The only estimate where the terms were jointly significant at the 5 percent level was for %ΔONE for 1929–1939. The estimated regression (standard errors in parentheses) was: