Opportunity and Overrides: The Effect of Institutional Public Support on Congressional Overrides of Supreme Court Decisions

Abstract

Existing theories of legislative-judicial relations emphasize the role of public support for the judiciary on the likelihood of legislative compliance. Although Congress can strengthen or weaken the Supreme Court’s decisions after initial compliance, the role of public support for the judiciary on subsequent legislative action is unclear. We develop a theory of legislative-judicial interactions, which suggests that Congress considers the court’s current level of public support when determining whether to override a Supreme Court decision. We test our theory using data on congressional overrides of US Supreme Court decisions, finding that high levels of public support for the court shield the court from hostile congressional action. The results underscore the vital role played by the public in interbranch relations, suggesting that public support plays a role in the legacy of a judicial decision beyond ensuring initial compliance.

Keywords

judicial politics interbranch relations public support congressional override

Public and scholarly interest in the public’s support for the US Supreme Court has blossomed in recent years. Journalistic accounts of the court’s support have suggested that the court’s support at the beginning of this decade reached perilously low levels (e.g., Pew Research Center 2013). Gallup reported in 2014 that only 30 percent of Americans had a “great deal” or “quite a lot” of confidence in the institution (Riffkin 2014). This low level of public confidence marked a historic low for the judiciary (McCarthy 2014).

Given the wealth of social scientific evidence emphasizing the vital nature of public support to institutions (e.g., Easton 1975; Caldeira and Gibson 1992), these reports suggest that the court’s position as an equal partner in governance may be in grave danger should these levels of support continue (cf. Bartels and Johnston 2013; Gibson and Nelson 2015). After all, courts rely on other actors to implement their decisions, and low public support increases the likelihood of legislative noncompliance, making it less likely that the court’s decisions are implemented (Vanberg 2005). The legislative electoral connection provides the mechanism: if legislators fail to implement the decisions of a popular court, they may suffer harmful electoral consequences; as the court’s support declines, so do the costs of legislative noncompliance (Carrubba 2009; Stephenson 2004). Hence, because both the court lacks implementation power and public support affects legislative compliance, low levels of public support put the judiciary at risk of impotency.

Yet, the court’s decisions are open to renegotiation long after initial compliance is achieved. Once a high court issues a decision, that ruling represents the state of the law until either the legislature or that same court takes actions that overrule that decision. Even without nullifying a decision, the legislature can continue to respond to the decision, subsequently taking actions that either strengthen or weaken that opinion’s legacy (Barnes 2004; Staudt, Lindstädt, and O’Connor 2007). While congressional responses to US Supreme Court decisions are fairly rare—less than 5 percent of US Supreme Court decisions are eventually overridden—these responses to Supreme Court decisions are substantively important. In the last two decades, Congress has used this power to negate US Supreme Court decisions on issues ranging from civil rights, intellectual property, illegal immigration, and equal pay for women (Christiansen and Eskridge 2014). As a result, understanding the circumstances that lead Congress to move the status quo away from the policy made by the US Supreme Court is essential to our understanding of interbranch relations in the United States.

What explains these congressional responses to US Supreme Court decisions? Surprisingly, the bulk of the evidence suggests that Congress’s ideological disagreements with precedents rarely affect congressional responses to Supreme Court decisions (Hettinger and Zorn 2005, but see Uribe, Spriggs, and Hansford 2014). The lack of an obvious role for ideology in this process suggests that other forces play a primary role. Given the important role that it plays immediately after the court decides a case, public support is an obvious candidate to explain the timing of subsequent congressional responses to the court’s decisions. Yet, though scholars have acknowledged that the public’s attitude toward specific decisions may affect congressional behavior (Ignagni and Meernik 1994; Ignagni, Meernik, and King 1998), to date, no systematic study has addressed the role of the court’s support on subsequent congressional actions toward the court’s decisions.

This omission is particularly surprising given the bevy of more general evidence that public support serves as a shield for political institutions, particularly courts, that protects their decisions from reprisal (Caldeira and Gibson 1992; Gibson and Caldeira 2009). This evidence leads us to theorize that public support affects subsequent congressional responses, just as it affects initial legislative compliance. Should these theories apply to congressional responses to judicial decisions beyond implementation, it would imply that the current historically low levels of public support for the court have even wider-ranging implications that put the court’s past decisions in peril.

In this paper, we look beyond initial compliance and examine the effect of the judiciary’s public support on Congress’s subsequent responses to judicial decisions. In contrast with previous studies (e.g., T. S. Clark 2011), which focus on the relationship between public support and judicial decision-making, we develop a theory of legislative-judicial interactions that underscores the importance of the court’s public support for congressional decisions. We argue that Congress acts strategically when considering when to respond to a judicial decision. Congress waits for the right opportunity to respond, based on the court’s level of public support. When the court’s support is low, Congress should be more likely to act in ways that overrule or limit judicial decisions because of a similarly low likelihood of a negative public response.

This approach posits a new avenue of influence of the court’s public support: policy implications. Overrides serve as more than simply court-curbing devices; they have real policy implications. While initial compliance and judicial self-restraint have some subsidiary policy implications, the primary purpose for overrides of Supreme Court decisions is to change the location of policy. The evidence we find suggesting that public support affects the likelihood of overrides is evidence that the court’s support can hurt more than just the legitimacy of the court; it also limits the court’s position in the policy-making process.

We test our theory by examining congressional overrides of US Supreme Court decisions using information about public support for the court over four decades. We find support for our theory. Our results suggest that when the US Supreme Court’s support is thriving, Congress is less likely to override the court’s decisions.

These results—particularly in a day and age when the media suggests that the court’s public support is near historic lows—suggest that public esteem for the court plays an even more important role in our political system than has been previously recognized. The role of public support reaches beyond compliance and affects subsequent legislative responses, as well. Just as low public support for the court can make Congress unwilling to comply with judicial decisions, it also can leave the court’s body of precedent open to congressional renegotiation.

The Roles of Institutional Public Support

Theories of the policy process are often based on the postulate that political actors make policy decisions based on their ideological preferences (Krehbiel 1998; Segal and Spaeth 2002). These theories apply both within and across institutions, suggesting that, when one branch of government in a system of separated powers passes a policy disagreeable to another branch of government, the second institution will nullify it. In the case of the US Supreme Court’s statutory decisions, Congress can pass a law that overrides a judicial decision, thereby nullifying it. Scholars have long been interested in the timing and effects of these overrides (e.g., Barnes 2004; Eskridge 1991a; Ignagni, Meernik, and King 1998). Conversely, ideology also acts as a preemptive constraint. Because their members are motivated by policy goals, institutions are unlikely to pass policies they know are likely to be nullified by the other branches (Epstein, Knight, and Martin 2001).

Yet, one of the most glaring empirical puzzles in separation of powers scholarship is the paucity of evidence supporting the role of ideology in interactions between Congress and the Supreme Court. Indeed, the lack of evidence supporting Marks’s (1989) canonical theory, which suggests a rational Supreme Court should be sensitive to the preferences of the other branches of government, has confounded scholars for a quarter century (Owens 2010; Segal 1997).¹

In the case of the US Supreme Court’s statutory decisions, Congress can pass a law that overrides a judicial decision, thereby nullifying it and moving the location of the status quo to a new location. Scholars have long been interested in the timing and effects of these overrides (e.g., Barnes 2004; Eskridge 1991a; Ignagni, Meernik, and King 1998). However, despite the strong theoretical (and intuitive) basis underlying the hypothesis that Congress is more likely to override decisions it disagrees with, scholars have again mostly failed to find evidence that congressional overrides are driven by ideological disagreement (Hettinger and Zorn 2005; Ignagni, Meernik, and King 1998).²

One likely reason for the lack of an ideological explanation for congressional overrides comes from Christiansen and Eskridge (2014), who examine congressional behavior between 1967 and 2011. Examining the circumstances that lead Congress to override a judicial decision, they note that a majority of congressional overrides do not occur in extremely salient policies—such as voting rights—which are the most commonly used as examples of such overrides. Rather, they (2014, 1370) write that about two-thirds of overrides involve Congress updating policies set by the court with ones “that Congress considered more equitable, more efficient, more consistent with current political values, or better suited to changed circumstances.” In this way, most congressional overrides represent a bipartisan desire by Congress to update a policy, albeit away from the status quo policy set by the court. The bipartisan nature of these overrides suggests that the direction of such changes may not be in the direction typically predicted by our ideological theories.

If ideology is not the driving force behind Congress’s decisions to move policy away from status quo policies set by the court, what else might explain congressional action? Previous literature has identified at least one possible explanation: invitations from the Supreme Court. Indeed, in some cases—about 20 percent by Christiansen and Eskridge’s (2014, 1370) count—Congress responds directly to an invitation by the court to override. However, the percentage of cases in which “responding to Supreme Court concerns” is the main reason for an override is far less than the percentage of overrides that “correct . . . a bad interpretation of the law,” “respond . . . to confusion in the law,” or “updat[e] policy.” In addition, it is not clear whether the Supreme Court includes invitations with the sole intent of signaling to Congress that they should override the decision. Rice (2015) shows that most invitations to override are not intended to bring about override, but are incorporated to indicate to lower courts that the issue belongs with Congress and is not subject to subsequent review by the courts.

We suggest a different mechanism through which Congress might choose to respond to a judicial decision at a particular point in time. A prominent literature presents a wealth of evidence that public support plays a key role in how institutions behave (see, for example, T. S. Clark 2011; Vanberg 2005). As applied to the relationship between the legislature and the judiciary, this literature has focused almost exclusively on how support for the judiciary affects either a court’s own decision-making tendencies or the legislature’s willingness to implement a given judicial decision. This is easily seen by reviewing three major findings.

First, the central body of scholarship on public support for the judiciary suggests that diffuse public support (more commonly termed legitimacy) is essential for an independent judiciary, because it enables political institutions to make unpopular decisions without popular reprisal (Caldeira and Gibson 1992). As Gibson and Caldeira (2009, 4) put it, “no political institution could be effective without some mechanism to believe that accepting their policy outputs, even disagreeable ones, is the right thing to do” (see also Easton 1975). By this account, public support affects interbranch relations by cushioning popular blowback caused by unpopular decisions. Thus, low public support opens the court to popular reprisal, but not necessarily to nullification of its policies.

Second, public support also affects institutions by promoting institutional self-restraint. T. S. Clark’s (2009, 3-4) account of the role of public support suggests that public esteem matters primarily by creating opportunities for members of Congress to engage in position-taking activities—the introduction of court-curbing legislation—that have little chance of becoming law but act as signals of low public support for the court. Thus, “because the Court relies on political will to give effect to its decisions, and because political will is often directed by public opinion, the most relevant constraining force on judicial power is public support for the Court.” Clark finds that these bills do induce self-restraint among the court’s justices, making them less likely to exercise their power of judicial review (and, therefore, change policy) after Congress introduces a spate of court-curbing legislation. Thus, by Clark’s account, low levels of public support induce self-restraint among the justices that makes them less likely to change the status quo because they fear nonimplementation of their decisions. Hence, Clark’s view of the role of public support is court-centric: the effect of low public support, by Clark’s theory, is to make courts cautious when their support is low and to embolden them to nullify congressional policies when it is high.

Third, public support promotes compliance with individual judicial decisions. Courts usually depend on other institutions to implement their decisions, and public support affects this process. Specifically, legislatures are more likely to comply with judicial decisions when support for the court is high (Vanberg 2005). When support is high, the costs of legislative noncompliance are raised (Carrubba 2009; Stephenson 2004). These costs bear an electoral form; a legislature that fails to comply with a popular court risks public reprisal in the next election (Vanberg 2000, 2001).³ Thus, by this view, low levels of public support matter at the time a decision is issued, suggesting that high levels of public support are necessary to put an opinion into practice.

An Opportunity Theory

These central insights from the literature on public support for the court suggest that higher levels of public support affect both the court’s own decision-making and Congress’s implementation calculus; these existing theories emphasize the importance of public support for the court at the time that a decision is issued. The literature is relatively silent on consequences of public support for the court on Congress’s subsequent treatment of a decision. This omission is particularly surprising because judicial decisions have long-reaching consequences; judicial decisions continue to have weight until they are overridden by Congress or by a future court. If, as Caldeira and Gibson (1992) and Easton (1975) suggest, higher levels of public support can cushion the court against negative public reaction in the short term, might this shielding effect of public support also be applicable against the legislature in the long term?

We answer this question in the affirmative, suggesting that Congress behaves proactively as it pertains to the court’s public support as a result of the electoral connection. Recall that both T. S. Clark (2011) and Vanberg (2005) suggest that electoral motivations underpin legislative action vis-à-vis the judiciary. On this point, Ignagni and Meernik (1994); Ignagni, Meernik, and King (1998); Meernik and Ignagni (1995); and Meernik and Ignagni (1997) all provide evidence that electoral motivations underlie subsequent congressional responses, just as they affect the introduction of court-curbing bills and initial legislative compliance.

A dispute over equal pay for men and women provides a concrete example of this logic. In the 2008 election cycle, many Democrats campaigned against the US Supreme Court’s decision in Ledbetter v. Goodyear Tire & Rubber Co., a ruling dealing with equal pay for men and women (Holland 2007). At the start of the next term, following the Democratic takeover, Congress passed the Lilly Ledbetter Fair Pay Act of 2009, which overrode the court’s decision; the bill was the first one signed into law by Barack Obama. According to data from the General Social Survey (GSS), which we describe in detail below, the proportion of the public who expressed little confidence in the court rose in the years after the court decided Ledbetter and the time when the decision was overturned. Thus, as this example and the prior literature demonstrate, the legislature may initially implement a decision if the court was popular when it was issued and wait until the court’s support falls before overriding the decision. Put differently, because public support has a shielding effect, high levels of public support will protect the court’s decisions from congressional overrides even long after initial compliance is established, but a drop in popular support for the court leaves its decisions open to congressional nullification, because the potential electoral backlash that is so potent when the court’s support is high fades away as the court’s support falls.⁴ Thus,

Hypothesis 1: The probability of a congressional override of a Supreme Court decision is inversely related to the court’s level of public support.

Data

Our theory suggests that Congress is more likely to override Supreme Court decisions when the court’s public support is low. A test of this theory requires reliable and valid measures of (1) congressional overrides in a given year, (2) the court’s public support in that year, and (3) an appropriate battery of control variables.

Identifying Treatments of Court Decisions: The Outcome Variables

We have all congressional overrides from 1973 to 2010 from Christiansen and Eskridge (2014). Our data begin in 1973 because that is the year in which data on support for the court are first available, as explained below. With these data, we are able to examine our hypothesis.

We focus on overrides of Supreme Court decisions, consistent with prior literature, which has focused on Congress’s decision to override these cases (Hettinger and Zorn 2005; Ignagni, Meernik, and King 1998; Spiller and Tiller 1996). The outcome variable in many of these prior studies has been based on a measure pioneered by Eskridge (1991a). Eskridge (1991a) relied on committee reports printed in the US Code Congressional and Administrative News to determine whether a decision had been overridden. However, in recent years, these committee reports have become a less reliable source for counting overrides. To this end, Christiansen and Eskridge (2014) provide an updated methodology for counting overrides and provide evidence that the updated measure is more valid than the Eskridge (1991a) measure. We use their listing of overrides in each year as the basis for our outcome variable. The data contain 244 overrides in our time frame.⁵ Many scholars have emphasized the distinction between overrides in congressional and statutory cases (Blackstone 2013; Hettinger and Zorn 2005). We do not expect our theory to differ in constitutional versus statutory cases. Moreover, Uribe, Spriggs, and Hansford (2014) show that there is little difference in how Congress approaches these types of cases, at least in terms of ideological divergence. The majority of overrides included in our data are statutory. In all, 204 (84%) of the overrides identified by Christiansen and Eskridge (2014) occur in statutory cases.

Measuring Support for the Court

We also require a measure of public support for the court.⁶ The only available time series of public support for the US Supreme Court comes from the GSS, which has asked respondents most years since the early 1970s for their attitudes regarding their confidence in the US Supreme Court.⁷ This measure is “the most reliable and consistent way to capture public support for the Court” (T. S. Clark 2011, 125), though we acknowledge that it is a measure of institutional confidence rather than institutional legitimacy.⁸ However, because our concept of interest is general public support (which blends short- and longer-term evaluations of the Court), rather than institutional legitimacy specifically, the GSS measure is appropriate as Gibson, Caldeira, and Spence (2003) have shown. Our measure of Court Disapproval follows from T. S. Clark’s (2009) and is the percentage of GSS respondents espousing “hardly any” confidence in the US Supreme Court.⁹ This variable has a median of 14.73 percent and ranges from 10.94 percent to 20.71 percent.

Control Variables

While an ideological connection has not always been well supported in the literature, it would be entirely inappropriate not to control for the most prominent alternate explanation for congressional overrides: ideological disagreement.¹⁰ Spatial theories of policy making predict that Congress is more likely to override judicial decisions it disagrees with when the decision is located farther from its most preferred policy. We model our ideological distance measurement on the approach in Uribe, Spriggs, and Hansford (2014), the only large-N study to find support for ideological disagreement. To measure the location of the precedent, we rely on Judicial Common Space scores (Epstein et al. 2007), using the ideal point of the median of the majority coalition as the measure of the location of the judicial decision.¹¹ We assume a chamber median model, where the left and right most pivotal members of the gridlock interval are determined by the medians of the two chambers and the president (Krehbiel 1991, 1998).¹² The left pivot is the ideology of the three members (House median, Senate median, and president) whose ideology measure is the smallest, and the right pivot is the member with the largest ideology measure. We define the gridlock interval as the space between the left and right pivots. If the case is located between these pivot points, the case is said to be in the gridlock interval. Here, we depart from the approach taken in Uribe, Spriggs, and Hansford (2014). We choose to separate out the gridlock interval from ideological distance. Thus, if the case is located between the pivot points, Gridlock Interval is equal to 1, otherwise, it is equal to 0. We then measure Ideological Distance as the absolute value of the distance between the court decision and the closer of the left and right pivotal members.¹³ We include in our analysis both Gridlock Interval and Ideological Distance, as well as an interaction between the two, to control for any differences in the effect of ideological distance inside the gridlock interval, where theoretically it should have no effect, and outside the gridlock interval.

As discussed above, T. S. Clark (2009, 2011) has argued that interinstitutional attacks can serve an important position-taking function that is electorally beneficial to legislators. Thus, if we do not take the electoral calendar into account, we risk attributing congressional actions on the basis of their electoral calendar to our theory, which predicts congressional action on the basis of public support. Thus, we control for whether or not the year is an Election Year. We expect, following Clark’s logic, that overrides should be more common in election years. This variable has the added benefit of capturing heterogeneity between policy making in the first and second years of a Congress.

Likewise, when choosing which case to treat, not all cases are created equal. Thus, we control for case-level salience using an indicator of whether the case was on the front page of the New York Times the day after it was decided, following Epstein (2000).¹⁴

The prior literature on congressional overrides presents a bevy of case-specific explanations that predict which cases Congress is likely to treat. With these in mind, we control for the same concepts suggested by Hettinger and Zorn (2005) to account for case-level variation in the likelihood of a congressional response.¹⁵ We include five additional dichotomous variables: whether a congressional amicus brief was filed in the case, whether the United States was the losing party to the case at the US Supreme Court, whether there are multiple legal provisions in the case, whether the Supreme Court decision was unanimous, and whether the Supreme Court reversed the lower court decision.¹⁶ All of these variables were measured as described by Hettinger and Zorn (2005).

Because the public is such a key actor in our theory, we also include a measure of the public’s divergence from the court first pioneered by Durr, Martin, and Wolbrecht (2000) to account for variation in judicial approval that is due to policy agreement. As Durr, Martin, and Wolbrecht (2000) show, this divergence can influence the public’s view of the court. This divergence measure increases as the public and the court are more distant from one another.¹⁷ Because the translation of changes in divergence to changes in support will be lagged, we use a two-year moving average of the divergence score.

In addition, given the explosion of congressional polarization during this time period, which likely influences the likelihood of an override as well as the size of the gridlock interval, we also include polarization as a predictor in the model. We measure polarization as the distance between the Democratic and Republican party means using Nominal Three-Step Estimation (DW-NOMINATE) scores, as suggested by Poole.¹⁸ The last item that we control for is multiple treatments of the same case. These data contain some repeat overrides of cases after the initial override.¹⁹ As this indicates that the Supreme Court decision does not “die” as in a standard survival model, we keep the case in our dataset after every override, and we include a variable that controls for whether the case has been previously treated before a given year.

Method

The theory implies the importance of time.²⁰ Although Congress can override a case any time after it is decided, the theory predicts that treatments are more or less likely given levels of the public’s support for the court. We model if and when Congress overrides a given US Supreme Court decision. We employ a discrete time duration model, consistent with recent approaches to modeling congressional overrides (Uribe, Spriggs, and Hansford 2014).²¹ This approach allows us to model the hazard shape directly. Following Carter and Signorino (2010), we use a cubic polynomial approach to model the hazard shape. Carter and Signorino (2010) suggest that this parameterization strikes a balance between finding the best hazard shape and not overfitting these data. Because we take this discrete approach, our data consist of case-year dyads. The outcome variable is whether case i is overridden in year t. Because this outcome variable is binary, we use a logit model in our analysis.

Results

Table 1 provides the results of our analysis. Recall that our theory predicts that disapproval should be positively related to the likelihood of an override.

Table 1.

Override Discrete Time Duration Model.

	Model 1
Court disapproval	0.08*
Court disapproval	(0.03)
Public mood divergence	0.01
Public mood divergence	(0.00)
Gridlock interval	0.11
Gridlock interval	(0.27)
Ideological distance	−0.63
Ideological distance	(0.97)
Election year	1.27*
Election year	(0.16)
New York Times front page	−0.03
New York Times front page	(0.20)
Congressional amicus	0.36
Congressional amicus	(0.34)
Unanimous	−0.61*
Unanimous	(0.15)
Multiple issues	0.24
Multiple issues	(0.15)
US lost case	0.33
US lost case	(0.24)
Reverse lower court decisions	−0.04
Reverse lower court decisions	(0.13)
Polarization	0.58
Polarization	(0.52)
T	−0.28*
T	(0.04)
t ²	0.01*
t ²	(0.00)
t ³	−0.00
t ³	(0.00)
Previously overridden	0.80*
Previously overridden	(0.33)
Gridlock Interval × Ideological Distance	1.19
Gridlock Interval × Ideological Distance	(1.34)
Intercept	−6.93*
Intercept	(0.65)
N	217,270
AIC	3,097.80

The outcome variable is whether a case is overridden in a given year. AIC = Akaike Information Criterion.

Indicates significance at p < .05.

These results support our theory. According to these estimates, Congress is more likely to override a decision when disapproval of the court increases. In short, as the court’s level of support increases, the probability of an override decreases.

How large are these effects? Importantly, the likelihood that Congress overrides any judicial decision in any year is very low. With this in mind, Figure 1 shows the predicted probability of an override. As Court Disapproval increases, the probability of an override increases. While Figure 1 makes it seem as though the difference over the range of this variable is small, due to the low probability of an override in any year, the percentage increase in predicted probability across the range of Court Disapproval is substantial. The probability of an override increases by about 109 percent from the minimum to the maximum of the observed court disapproval scores.

Figure 1.

This graph shows the predicted probability of an override across values of the court’s public disapproval measure.

These results also underscore the importance of electoral motivations in the decision to override Supreme Court decisions. The probability of an override is substantially higher in an election year than in a nonelection year. An override is 256 percent more likely in an election year than in a nonelection year.

Some of the other estimates also merit discussion. Consistent with much of the previous literature, we do not find evidence of an ideological effect on the likelihood of override.²² We do not find evidence of an ideological effect using either our distance measure or our gridlock interval measure. The ideological distance variable does not reach conventional levels of significance either inside or outside the gridlock interval. We do find evidence that cases that have been previously overridden are substantially more likely to be overridden in the future. This result suggests that if Congress has responded to any portion of a case in the past, the rest of the case is more vulnerable to future overrides. Consistent with previous literature, we find that overrides of unanimous cases are less likely. We also find that the likelihood of override decreases exponentially as the case ages. The likelihood of override is highest in the first few years after the decision is passed.

Discussion and Conclusion

In his essential study of legislative-executive relations, T. S. Clark (2011, 4) argues that “elite will is not necessarily enough to check the courts; rather the separation of powers requires a degree of public will to ‘rein in’ the judiciary.” Our analysis takes this point one step further, suggesting that public support can restrain the legislature, as well as the judiciary. Our data indicate that Congress shows self-restraint by failing to negatively respond to judicial decisions when the court’s support is high. However, low levels of public support for the court signal to legislators that the electoral costs of nullifying judicial policies are negligible, enabling Congress to override a judicial decision without fear of popular reprisal.

The implications of the finding are both important and distinctive. First, the findings demonstrate that maintenance of high levels of public support in the long term is essential for the court. Just as theories of judicial legitimacy would suggest, high levels of public support act as a shield for the court’s policies. If the court’s public support declines in the present, the continued acceptance and implementation of its past decisions are at risk. In short, public support is not only important for the court’s short-term policy goals; it needs high levels of public support to maintain the policies espoused by its previous decisions.

Second, the focus on the effect of the court’s public support on Congress is distinctive. We demonstrate that the court’s public support has meaningful consequences for the long-term success of the policies it espouses in its opinions. Whereas previous theories of legislative-judicial relations (e.g., T. S. Clark 2011) have emphasized the court’s reactions to fluctuations in its own public support, we show that Congress also responds to changes in the court’s public support, becoming more likely to nullify the court’s policies when the court’s own public support is low.

Indeed, these results underscore the important role that the court’s public support plays in Congress. T. S. Clark’s (2011) theory suggests that members of Congress at least indirectly reflect the court’s public support in their patterns of bill introductions, but because those court-curbing bills rarely become law, there is no direct relationship between congressional policy making and public support for the court. Our theory expands the role of public support for the court, showing that Congress also responds to the court’s level of public support in the bills that it passes. Indeed, overrides are not merely symbolic acts like the introduction of court-curbing legislation. Rather, overrides change policy, nullifying a decision of the US Supreme Court. In short, these results suggest that the court’s support has more expansive effects than have heretofore been documented.

The effect of public support on the court is a substantively meaningful one. The likelihood that Congress will treat a given Supreme Court opinion in a particular year is very low, as is the probability that a country will transition to a democracy in a given year, that a state will fail in a given year, or that two countries will go to war in a particular year. Yet, all of these outcomes are politically important ones, affecting national security, the quality of governance, and the efficacy, in our case, of a constitutional system. The ability of Congress to check the court by overriding its decisions is a fundamental power of that institution; simply because it is rarely employed does not make it an unimportant outcome. Indeed, to reiterate a result from above, this study demonstrates that a massive change in the court’s public approval can make a congressional override of a particular decision more than 100 percent more likely. This is a substantively important finding.

Returning to the journalistic hand-wringing that opened the paper, our analyses, coupled with the findings of Clark and Vanberg, indicate that low public support has three negative consequences for the court: it makes the court less likely to act assertively to nullify congressional action, it increases the likelihood of legislative noncompliance, and it leaves the court’s past decisions open to congressional nullification. Hence, public support for the court affects both initial compliance as well as subsequent congressional actions regarding judicial decisions. Moreover, because of the policy-based nature of congressional overrides, our findings here show explicitly for the first time that the court’s support has real policy considerations in addition to the institutional legitimacy considerations shown in previous studies.

These findings underscore the vital nature of public support to the judicial branch of government. Lacking a direct electoral connection to periodically replenish its public support, the court is dependent upon public approval to fulfill its role as an equal partner in American governance and to ensure that, over time, its decisions continue to be respected by the legislature. Although we have known that public sentiment at the time of the decision affects the court’s decisions (Casillas, Enns, and Wohlfarth 2011; Epstein and Martin 2010) and the likelihood of implementation (Vanberg 2005), our study illuminates that, even after initial compliance is established, the court’s support still influences the longevity of its decisions; if its support dips, then its body of prior decisions are at risk of override.

Beyond the implications of these findings for the court, our results also shed light on the efficacy of the American system of separated institutions sharing power. Given the paucity with which previous research has found evidence supporting ideological separation-of-powers considerations in the Supreme Court’s decisions in the past (Owens 2010; Segal 1997), this suggests a new mechanism through which the court might consider Congress’s preferences. It could be that the court’s historically high levels of support help to explain why scholars have been unable to find evidence of separation of powers considerations in Supreme Court voting, with the court relying on its “reservoir of goodwill” to ward off the most severe attacks from Congress. Future research should further examine how public support for the court affects the relationship between Congress and the Supreme Court.

Likewise, future research may want to examine how variation in the level of public support for the court affects the success of a congressional override. The research design we have employed necessarily treats all overrides as equal, but the effects of overrides might differ in many ways: the extent to which they move policy, the extent to which they affect levels of judicial dissensus (a possibility ably documented by Barnes 2004), the amount of discretion they provide to implementing actors, and the extent to which the implementation of the override results in de facto differences in policy. It may be that public support has a second-order effect, only making an override less likely to happen but also making an observed override less effective in practice. This possibility is ripe for future research.

To return to the empirical observation that began the paper, what might the longer-term consequences of the court’s lower than typical levels of public support have on its policies? Our results suggest that, should the court’s support continue to decline, Congress will feel freer to override the court’s decisions, therefore, undermining the stability of the law. Moreover, should the court’s support continue to decline, an emboldened Congress would be able to disempower the court in the latter’s attempts to be a coequal partner in governance: when Congress is willing and able to override the court with impunity, the court’s ability to be an effective check on congressional and presidential power begins to waver. In a time where many are looking to the court to fulfill this constitutional role, our findings suggest that the court’s low public support render it unable to be the knight in shining armor wished for by many.

Footnotes

Acknowledgements

We thank Marcus Hendershot, Rachael Hinkle, Keith Schnakenberg, Morgan Hazelton, James Rodgers, and the faculty seminar at the University of Illinois for helpful feedback.

Authors’ Note

Previous versions of this paper were prepared for the 2014 meeting of the American Political Science Association and the 2015 meeting of the Western Political Science Association.

Declaration of Conflicting Interests

The author(s) declared no potential conflicts of interest with respect to the research, authorship, and/or publication of this article.

Funding

The author(s) received no financial support for the research, authorship, and/or publication of this article.

Supplemental Material

Replication data for this article is available at .

Notes

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