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Public Choice

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12.09.2016

Supermajority rule and bicameral bargaining

verfasst von: Dongwon Lee

Erschienen in: Public Choice | Ausgabe 1-2/2016

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Abstract

This paper revisits the claim that supermajority rules and bicameral legislative structures restrain excessive government spending and taxation. Our analysis suggests that the extension effect of a supermajority rule—that requires logrolling across additional members—increases with the ratio of seats in the House relative to seats in the Senate. Using a panel of US states, 1970–2008, we find that the ratio of House-to-Senate seats has a robust, positive impact on the tendency of a supermajority rule to inflate the budget. Our finding implies that a supermajority rule can have a perverse effect on budget outcomes in bicameral legislatures owing to two factors: the geographic overlap between chambers and the low price elasticity of demand for public goods.

Vorheriger Artikel Equitable representation in councils: theory and an application to the United Nations Security Council

Nächster Artikel Identity voting

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This argument assumes that the marginal cost of building a coalition increases with the size of the coalition. For instance, the cost of achieving a simple majority is higher in a 100-member legislature divided into 66 representatives and 34 senators than in a legislature equally divided between House and Senate—because the additional cost of buying eight more votes (34–26) in the House exceeds the saving from buying eight less votes (18–26) in the Senate (McCormick and Tollison 1981). We thank William F. Shughart II for pointing out this argument.

This assumes that House proposers cannot target large projects to their own districts.

In a dynamic setting, supermajority rule can increase majority tyranny because the ruling supermajority is more likely to avoid future retaliation by the current minority—who has a smaller chance of achieving a supermajority in the future (Dixit et al. 2000).

Although some studies analyzed the effect of supermajority rule in bicameral settings, the chambers were treated as independent unicameral legislatures with no veto powers. See Bradbury and Crain (2001) for a similar discussion in the context of the law of 1/n.

For instance, House-to-Senate seat ratio of 1 indicates a complete constituent homogeneity across chambers—that is, two chambers have equal bases of representation.

We assume that there is no spillover in project benefits across districts (to focus on the case of distributive goods). See Besley and Coate (2003) and Lockwood (2002) for a local public good model with a degree of spillovers.

We assume that \(n \ge 3\) and \(\tau _{j} \ge 1\) \(\forall j\).

Alternatively, projects can be targeted to a Senate district, and the benefits are divided equally among all House districts within the targeted district (e.g., Chen and Malhotra 2007). Our results do not change qualitatively under the alternative setup, although the current setup allows us to pinpoint the effects of supermajority rules in both chambers.

For instance, formally, \(\alpha n=\min \left\{ a\in N\mid a\ge \alpha n \right\}\) where N is the set of natural numbers.

Chen and Malhotra (2007) suggested that representatives usually do not work with representatives in different Senate districts—even with geographically close representatives—due to logistical and other transaction costs. Similarly, Persson et al. (1997) noted that bargaining within a chamber is less costly than bargaining across chambers.

That is, a proposal is made in stage 1, and legislators in both chambers vote on the proposal in stage 2.

Consistent with Ansolabehere et al. (2003), the marginal value of a senator is zero. The \(\alpha n \tau\) House districts are large enough to win the \(\alpha n\) Senate districts.

A House member’s continuation value C is the expected payoff in any round. Suppose that a representative is recognized with probability \(\sigma\), and a senator recognized with probability \(1 - \sigma\). Then, \(C = \sigma \big [ \frac{\alpha n \tau - 1}{n \tau } \times C + \frac{1}{n \tau } \times (G - (\alpha n \tau - 1)C) + \frac{n \tau - \alpha n \tau }{n \tau } \times 0 \big ] + (1 - \sigma ) \big [ \frac{\alpha n \tau - b_{k}}{n \tau } \times C + \frac{b_{k}}{n \tau } \times \frac{G - (\alpha n \tau - b_{k})C}{b_{k}} + \frac{n \tau - \alpha n \tau }{n \tau } \times 0 \big ]\). This indicates that \(C = G/n \tau\).

This result is similar to the budget outcomes under a supermajority rule (Lee et al. 2014; Lee 2015).

For instance, a simple majority can exploit the tax base of 49 % of the polity, while a 3/4 majority rule limits the tax base to just 25 % of the polity (Bradbury and Johnson 2006).

See Borcherding (1985), Reiter and Weichenrieder (1997), Brasington (2002), and Oates (1996, 2006) for the estimation of the price elasticity of demand for government services.

In some previous models (e.g., Chen and Malhotra 2007), an increase in House-to-Senate seat ratio reduces the incentive of lower chamber proposers to pursue large projects. In our model, the agenda setter keeps the bargaining surplus, effectively sharing in the projects of other coalition members. Thus the agenda setter proposes the spending projects of size G [in Eq. (1)]—even if an increase in House-to-Senate seat ratio reduces the agenda setter’s payoff from her own district. In addition, our paper focuses on the interaction between the supermajority rule and the ratio of House-to-Senate seats, in which the coalition success rule \(\alpha\) varies rather than being fixed at a simple majority.

For instance, in the U.S. Congress and other federal systems, the Senate represents geographical areas (states), the size of which rarely changes. At the U.S. state level, few state legislatures have reduced the number of upper house districts (n) without also changing the size of the lower house (T).

Nebraska has a unicameral legislature.

Other bicameral legislatures with supermajority rules include the European Union and the democratic countries that require a supermajority to override an executive veto.

Out of 17 states with a supermajority rule, 15 states adopted the rule after 1970.

During the sample period, 21 states changed the size of legislature, either House or Senate, at least once.

Most of the socioeconomic data were obtained from the U.S. Census Bureau; the income per capita data were collected from the U.S. Bureau of Economic Analysis. Legislative votes for constitutional amendment (one of our instrumental variables) were collected from the Book of the States.

We ignore the potential multicollinearity between \(S_{it}\) and \(Ratio_{it}\) because the correlation between the two variables is low (\(-0.13\)).

The neighborhood adoption dummy is one if neighboring states have adopted a supermajority rule within 10 years.

Although some states have amended their constitutions to alter the legislative size, these changes were motivated by all legislative matters, including efficiency in representation and other exogenous events (banking crises, political scandals, court decisions regarding equal population reapportionment, and statewide legislative reforms), not limited to budget allocation (Lee 2015).

More specifically, S is statistically significant in columns 4 and 6 (with both \(S*Ratio\) and \(S*Upper\) included), but insignificant in columns 3 and 5 (with only \(S*Ratio\) included). With interaction terms included in a regression, however, the coefficients on S alone becomes less meaningful. In fact, it is possible that the marginal effects of S are significant even if the point estimates on S are insignificant (Brambor et al. 2006).

Another potential reason for the inconsistent effects of Ln(POP) is the use of natural log form. In any case, our main results are qualitatively similar when Ln(POP) is dropped.

The first stage regressions reported in Appendix Table 8 show that the coefficients on the neighboring state effects and the constitutional amendment rules are statistically significant in most cases.

Most coefficients on the key variables (S, \(S*Ratio\), \(S*Upper\)) are significant in columns (1) through (5), but results are mixed in columns (6) through (8). This explains that the marginal effects of S are generally insignificant in the latter cases (i.e., columns (6) and (8)).

Proposition 39 reduced the required vote on local school bond measures from two-thirds to 55 percent; Proposition 25 reduced the vote requirement to pass the budget from two-thirds to a simple majority.

At the state level, the House-to-Senate seat ratio equals the average Senate district fragmentation only if House districts are completely nested within the Senate district.

Note that a Senate proposer obtains a project for only one of her House colleagues. The proposer can distribute surplus from the bargaining to the other House districts within her own Senate territory.

Thus, any one House district that overlaps with the Senate district j has \(1/a_{j}\)th of the Senate j’s population.

By standard assumption, u is concave and increasing in \(x_{i}\).

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Titel: Supermajority rule and bicameral bargaining
verfasst von: Dongwon Lee
Publikationsdatum: 12.09.2016
Verlag: Springer US
Erschienen in: Public Choice / Ausgabe 1-2/2016
Print ISSN: 0048-5829
Elektronische ISSN: 1573-7101
DOI: https://doi.org/10.1007/s11127-016-0369-9

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