MEASURING THE INCIDENCE OF ENDOGENOUS POLICIES: APPLICATIONS IN FISCAL FEDERALISM

National Tax Association

MEASURING THE INCIDENCE OF ENDOGENOUS POLICIES: APPLICATIONS IN FISCALFEDERALISMAuthor(s): Brian KnightSource: Proceedings. Annual Conference on Taxation and Minutes of the Annual Meeting ofthe National Tax Association, Vol. 94 (2001), pp. 357-362Published by: National Tax AssociationStable URL: http://www.jstor.org/stable/41954740 .

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MEASURING THE INCIDENCE OF ENDOGENOUS POLICIES: APPLICATIONS IN FISCAL FEDERALISM

Brian Knight , Board of Governors of the Federal Reserve System*

TO compare dence, MEASURE

of economic public THE ECONOMIC

policies, outcomes

EFFECTS, analysts across

OR

states often INCI-

dence, of public policies, analysts often compare economic outcomes across states

with and without such policies. However, their analyses of policies and economic effects ignore the role of state characteristics, such as preferences for public goods, which may influence both economic outcomes and the decision by states of whether or not to adopt such policies (Besley and Case, 2000). Given that state characteristics are often unobserved, statistical correlations between policies and economic outcomes may reflect the role of these state characteristics rather than the effect of policies themselves.

This policy endogeneity, a correlation between policies and state characteristics, may explain three longstanding empirical findings in public finance that do not reconcile with economic theory. First, intergovernmental grants do not crowd out spending by recipient governments. Second, fiscal restraints have little effect on budgetary outcomes. Third, capital taxes do not affect the location of investment.

This paper uses lessons from theoretical models of political economy to correct for the endogeneity of these three policies: intergovernmental grants, fiscal restraints, and capital taxes. The political economy models are used first to formally document the bias induced by correlations between policies and state characteristics. To correct for this bias, these models also provide a framework for selecting instruments based on procedures that govern the adoption of policies. These instruments separate the variation in policies into two components: one component due to characteristics of states and one due to the procedures governing the adoption of policies. Using only the latter source of variation, which is arguably independent of state characteristics, one can isolate the true economic incidence of policies.

*This paper is based on my University of Wisconsin-Madison dissertation. I am indebted to my advisers, Arik Levinson, Bob Haveman, and Andy Reschovsky, for their support and guidance, and to the Christensen Award in Empirical Economics and the Wisconsin Alumni Research Foundation for financial support. The views presented are solely mine and do not necessarily represent those of the Federal Reserve Board or its staff.

INTERGOVERNMENTAL GRANTS AND CROWD-OUT

According to some economic models, intergovernmental grants crowd out state spending on public goods, leading to little or no increase in combined public spending (Bradford and Oates, 1971a and 1971b).1 The existing empirical literature has found that grant receipts do not crowd out state spending.2 This empirical literature has assumed that the distribution of grants is independent of state characteristics. Such grants, however, are determined through a bargaining process between representatives of these states and may thus be endogenous, or correlated with state preferences, implying a bias in the existing literature against measuring crowd-out. Accounting for this correlation, estimators that use instruments for grant receipts document crowd-out that is economically and statistically significant (Knight, 2002). Crowd- out predictions are based on a model with an exogenous distribution of grants, as studied in Bradford and Oates. A single state allocates re- sources between a public good (G) and a private good (z). The state has N identical residents, each with a Stone-Geary utility function:

(1) U(G,zii) = ßH(G- ß)!P)] + (1 - ß)'n[z]

where ļi is the minimum public spending parameter, P is the relative price of public goods, and ß is the utility weight on consumption of public goods. The public good is financed through either federal grants-in-aid (A), which are given exog- enously in this section, or taxation of community income (M). Thus, the recipient state faces the fol- lowing resource constraint:

(2) PG + NZ¡=A+M.

Maximizing (1) subject to (2) yields a state spending function:

(3) g = ßM+(ß-')A + (l-ß)n

where g is public spending by the state, a measure which is net of federal grant receipts. As shown in

357



equation 3, this simple model predicts that federal grants crowd out spending by recipient states dollar for dollar, after accounting for income effects o?).

To test this crowd-out prediction, I study the effects of the Federal Highway Aid Program on highway spending by state governments using data from both the Census Bureau Census of Governments and the Federal Highway Administration (FHA) Highway Statistics Series. As displayed in the first two columns of Table 1, the ordinary least- squares (OLS) estimates from both data sources find little evidence of such crowd-out.

This rejection of crowd-out, in both Table 1 and existing empirical literature, relies on an assumption of exogenous grants. However, such grants are determined through a bargaining process be-

NATIONAL TAX ASSOCIATION PROCEEDINGS

tween representatives of these states and may thus be endogenous. To formalize this endogeneity ar- gument, I construct a legislative bargaining model that, besides studying the crowd-out effects de- scribed above, incorporates the political process governing the allocation of grants across states.3 A majority-rule legislature, consisting of one representative from each state, bargains over the distribution of federal grants. States differ in their preferences for public goods, as reflected in the Stone-Geary parameter (jn ). A committee chair, to whom significant agenda-setting powers are granted, constructs a minimum winning coalition consisting of those representatives whose votes are relatively cheap to secure. In the legislative bargaining model, votes of representatives from states with a strong preference for public goods are

Table 1 Federal Grant Crowd-out, Selected Coefficients

(** denotes 5% significance, * denotes 10% significance) (1) (2) (3) (4)

Estimator OLS OLS 2SLS 2 S LS Data Source Census FHA Census FHA Primary Coefficients Per-capita grant receipts -0.0327 0.1361 -0.8786 -0.9099

(0.0559) (0.0573)** (0.4199)** (0.4023)**

Per-capita income after federal taxes 0.0094 0.0087 0.0100 0.0086 (0.0011)** (0.0011)** (0.0013)** (0.0013)**

U.S. House Instruments transportation committee -3.9545 -3.7611

(9.3307) (8.9289)

majority party -7.8995 -7.5157 (6.4774) (6.2002)

tenure -0.9621 -1.2125 (0.4265)** (0.4101)**

U.S. Senate Instruments transportation committee 9.1975 1.4053

(6.2833) (6.0137)

majority party 1.1518 3.8492 (2.9091) (2.7851)

tenure 0.9268 1.0868 (0.2954)** (0.2828)**

Notes: 1 . Per-capita state government highway spending, net of federal grants, is the dependent variable in each regression. Additional regressors, not reported here, include state population, drivers per capita, vehicles per capita, gubernatorial/legislative partisan representation, and state fixed effects. Political instruments are measured as av- erages across the delegation. FHA refers to Federal Highway Administration.

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cheaper to secure than the votes of representatives from states with a weak preference. This differen- tial cost of securing votes induces a positive correlation between preferences for public goods and grant receipts (pA^> 0).

This theoretical correlation has implications for the existing empirical literature. Using equation 3, the probability limit of the grants coefficient ( bA ), the crowd-out measure in typical regressions, is given as follows:

(4) plim(^) = (ß - 1 ) + ( 1 - ß){aßlaA)pAit .

Thus, the existing empirical literature, which typi- cally assumes an exogenous distribution of grants, is biased against measuring crowd-out because of a positive correlation between grant receipts and preferences for public goods.4

To correct for this bias, I employ a two-stage least-squares estimator that uses instruments motivated by the legislative bargaining model in the theoretical section. In this model, the committee chair uses proposal power to secure above- average grants for his home state. This proposal power advantage suggests using instruments based on representation on transportation authorization committees, which have legislative authority over federal highway grants. Interpreting such proposal power more broadly, I also include as instruments both tenure and affiliation with the majority party. The last two columns of Table 1 show that political power within Senate delegations increases grant receipts.5

Using this exogenous variation in grant receipts, an additional dollar of grants reduces state spending by roughly 90 cents, as reported in Table 1. In contrast to traditional regression methods in this paper and the literature, the endogeneity-corrected results demonstrate that federal grants crowd out state spending in an economically and statistically significant manner, and this evidence supports the theoretical predictions of Bradford and Oates.

SUPERMAJORITY VOTING REQUIREMENTS AND TAXES

Traditional public finance research has tended to ignore the role of fiscal restraints, which are budget rules designed to place procedural hurdles in the way of tax or spending increases. Recent research has shown that a particular type of fiscal restraint, supermajority voting requirements for tax

increases, plays an important role in explaining differences in state budgetary outcomes (Knight, 2000).

A key issue in measuring the effects of fiscal restraints is institutional endogeneity. States endog- enously adopt these restraints, and may do so for reasons related to tax policy. As an empirical dem- onstration of this endogeneity, the first column of Table 1 provides the results of an OLS regression of effective tax rates on supermajority requirements, which are measured as the legislative vote required to raise taxes.6 This result suggests that supermajority requirements do not reduce taxes, since this coefficient is small and statistically in- significant.7 It is difficult, however, to rule out alternative explanations based on the propensity of states with certain attitudes toward taxation to adopt supermajority requirements.

To formally document this policy endogeneity, I first construct a two-stage legislative bargaining model that incorporates both supermajority requirements for legislated tax increases (the constitutional stage) and tax rates (the resulting statutory stage). The model demonstrates the counterintuitive result that pro-tax states are more inclined to enact supermajority requirements. Legislatures use these rules to reduce the agenda-setting power of pro- tax party extremists in the statutory stage. In sup- porting this requirement, the legislator with the ideal rate that is the median among all legislators is willing to give up being the pivotal voter in return for a more moderate policy outcome, namely the ideal rate.

To eliminate the bias induced by this endogeneity, the constitutional stage of the theoretical model motivates using the difficulty of amending state constitutions as an instrument for supermajority rules. Particular measures include the legislative vote required to initiate amendments, the number of legislative sessions required to consider amendments, and voter access to amendments through direct legislation. The first-stage coefficients in Table 2 all have the expected sign; states with constitutions that can be more easily changed are more likely to adopt supermajority requirements.

Using this source of exogenous variation in supermajority requirements, the second column demonstrates that the requirements reduce state tax rates in an economically and statistically significant manner.8 The tendency for pro-tax states to adopt supermajority requirements masks the true effect of such requirements. Traditional public fi-

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Table 2 Determinants of State Tax Rates,

Selected Coefficients (* * denotes 5% significance, *denotes 1 0% significance)

(1) (2) Estimator O LS 2SLS Primary Coefficient Supermajority Percentage -0.0057 -0.2136**

(0.0102) (0.0509)

Instruments Direct Legislation Indicator 0.0202**

(0.0033)

Legislative Vote Required to -0.0298 Amend Constitution (0.0184)

Sessions Required to -0.0120** Amend Constitution (0.0033)

Notes : Additional regressors, not reported here, include indi- cators for traditional tax and expenditure limitations, state income, grant receipts, and gubernatorial/legislative partisan representation. Tax data are taken from the Census Bureau Census of Governments.

nance research has tended to ignore such fiscal institutions, but this paper finds that supermajority requirements play an important role in explaining budgetary outcomes.

CAPITAL TAXES AND THE LOCATION OF INVESTMENT

The empirical literature on capital taxes has found little or no relationship between capital taxes and the location of investment.9 The literature typi- cally assumes exogenous taxes, however, and this weak relationship may reflect policy endogeneity, a propensity of states with favorable location characteristics to set high tax rates. Recent research that has corrected for such endogeneity has documented an economically and statistically significant relationship between capital taxes and the location of investment (Knight, 2001).

To account for this endogeneity, I first construct a tax competition model based on Zodrow and Mieszkowki (1986). 10 In the model, capital owners choose to invest among S states, which tax investment to finance a public good (G). To produce a private good, a perfectly competitive firm in each state has access to a decreasing returns technology f(k), where k is the ratio of capital to labor in the state.


Capital owners invest in the state that maximizes after-tax returns to capital ļfk- r], where fk is the marginal product of capital and ris the unit tax on capital. In equilibrium, a no arbitrage condition of equal rates of return holds across all states:

(5) ft-T=Ķ,

where R0 is the rate of return to an outside investment option.

A representative constituent, assumed immobile, gains utility from consuming both the public good and a private good (z), which is financed through after-tax returns to capital and labor. The fiscal choices in each state, designed to maximize the welfare of the representative constituent, can be expressed as a modified Samuelson condition:

(6) L*MRS = c/( l+£)

where L is state population, MRS is the marginal rate of substitution between public and private goods, c is the unit cost of producing public goods, and £ is the elasticity of investment with respect to capital taxes.

The condition of equal rates of return in equation 5 forms the basis for the estimating equation, while the modified Samuelson condition in equation 6 demonstrates the endogeneity of tax rates and motivates instruments to correct for such endogeneity. For empirical purposes, consider a linear parameterization for state productivity:

(7) fk = a[Xfi+Ç-k]

where X and £ are observed and unobservable location characteristics. Inserting equation 7 into equation 5 yields a regression equation:

(8) k = Xß+£-(l/a)r-RJa.

To measure the responsiveness of investment to taxes, the empirical analysis uses state-specific data on per- capita manufacturing capital and investment from the Census of Manufactures in 1977, 1982, and 1987." As shown in Table 3, neither the flow of investment nor the stock of capital responds to capital taxes in the traditional regression results.12 As demonstrated in the theoretical model, however, capital taxes are endogenous because they are set by states competing for investment. That is, tax rates may be correlated with unobservable state

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94™ ANNUAL CONFERENCE ON TAXATION

Table 3 Elasticity Estimates:

Taxes, Investment and Capital (** denotes 5% significance , * denotes 10% significance)

-1 -2 Estimator O LS 2SLS Manufacturing Capital 0.09* 0.28** Manufacturing Investment 0.23** 0.62** Notes: Additional regressors, not reported here, include state electricity prices, natural gas prices, land prices, percent unionized, road mileage, and investment market dummy variables for both 1977 and 1982. Instruments in 2SLS analysis include state population, percent of population under age 18, percent over age 65, percent in metro area, population density, square miles, gubernatorial/legislative partisan representation, and characteristics of competitor states.

characteristics (£), and this correlation may explain the weak results in both this paper and the existing literature.

To correct for this endogeneity, the tax competition model suggests instruments based on state population (L), public good production costs (C), preferences for public goods (MRS), and characteristics of competitor states, which are captured in the investment elasticity (£). Using proxies for these theoretical variables, the two-stage least squares (2SLS) results in Table 3 suggest larger elasticities. Thus, a correlation between capital tax rates and unobservable location characteristics may mask the true deterrent effect of taxes on manufacturing investment. In contrast to the existing literature, the endogeneity-corrected estimates in this study suggest that capital taxes do affect the distribution of investment across states.

CONCLUSION Policy endogeneity is an important complica-

tion in empirical analyses of the incidence of policies. This paper provides a framework for ad- dressing and correcting this problem by theoreti- cally and empirically incorporating the procedures of policy adoption. The approach seems particu- larly promising given recent advances in formal modeling of political processes.13 Although this paper has focused on issues in fiscal federalism, the methodology has potential for application to a wide variety of questions in public finance and economics.

Notes 1 Upon receiving grants, states redistribute the funds to constituents through tax cuts.

2 For recent reviews of this literature, see Hines and Thaler (1995) and Oates (1999). This finding has be- come known as the flypaper effect; that is, intergovernmental grants "stick" in the public sector.

3 This legislative bargaining model is based on Baron and Ferejohn (1987, 1989). 4 This expression is derived using an assumption of non- redi stribu tive grants (pAM = 0). 5 The House instruments have a counterintuitive nega- tive sign. There are two possible explanations for this divergence. First, the Democratic Party controlled the House, but not the Senate, for the entire sample pe- riod, providing little time-series variation in these House instruments. Second, the area represented by senators (the state) but not the area represented by House representatives (the congressional district) matches the unit of observation in the empirical model.

6 For states without supermajority requirements, the supermajority percentage is simply set at 50 percent. 7 According to the point estimates, adoption of a two- thirds supermajority requirement lowers effective tax rates by 01 percentage points, a small decrease relative to the average effective tax rate of 7.0 percent. 8 Introduction of a two-thirds supermajority requirement reduces effective tax rates by 3.6 percentage points, a large effect relative to the average effective tax rate of 7.1 percent. However, alternative two-stage least squares results, not reported here, suggest that a two- thirds supermajority requirement reduces tax rates a smaller amount, between 0.6 and 1 .6 percentage points. 9 See Wasylenko (1997) for a recent survey of this literature.

10 See Wilson (1999) for a recent survey of the theoretical tax-competition literature.

1 1 Each year is treated as a separate investment market; year dummy variables control for differences over time in returns to the outside option (R0). 12 Because of its static nature, the tax competition model does not inform as to whether the flow of investment or the stock of capital is the appropriate dependent variable. Therefore, I present results using both measures.

13 See Persson and Tabellini (2000) for an overview of the theoretical literature.

References Baron, David, and John Ferejohn.

Bargaining and Agenda Formation in Legislatures. American Economic Review 77, 2 (1987): 303- 309.

Bargaining in Legislatures. American Political Science Review 83, 4 (1989): 1 181-1206.

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to Collective Fiscal Decisions. Quarterly Journal of Economics 85, 3 (1971): 416-439. (1971a)

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MEASURING THE INCIDENCE OF ENDOGENOUS POLICIES: APPLICATIONS IN FISCAL FEDERALISM