U.S. District Court Decision-Making in USA PATRIOT Act Cases after September 11

Abstract

Our analysis of federal district court decisions in USA PATRIOT Act cases after September 11 shows that while district courts exhibit great deference to the executive during times of war, especially in terrorism threat and immigration cases, they are not inclined to protect civil rights and liberties during times of domestic or international conflict. While judicial ideology does not influence case outcomes, the presence of a civil-libertarian interest group decreases the chance of a deferential decision, and a more conservative Senate and president increases the chance of judicial deference.

KEYWORDS::

• USA PATRIOT Act
• terrorism
• judicial decision-making
• federal district court

Notes

While he discovers that the ideological preferences of individual circuit court judges influenced deference decisions, somewhat paradoxically Sunstein (2008:7) also reports that ideology did not influence decisions made by three-judge panels in circuit court cases; that is, the voting decisions of Republican (or Democratic appointees) are not affected by how many Republican (or Democratic appointees) are sitting on the panel deciding the case.

Typically, civil cases involve diverse set of claims, including foreign tax liability, racketeering, and cigarette smuggling (European Community v. Japan Tobacco, Inc., 2002), tort and contract liability in agreements for medical supplies (Medical Supply Chain, Inc. v. US Bancorp, 2003), regulatory jurisdiction over telemarketing fraud (National Federation of the Blind v. F.T.C., 2004), and the loss of trade secrets by virtue of computer virus infection (Resdev, LLC v. Lot Builders, Ass’n, Inc., 2005).

“American Civil Liberties Union, “Keep America Safe and Free,” http://www.aclu.org/keep-america-safe-free, accessed June 21, 2012.

Electronic Frontier Foundation, https://www.eff.org/, accessed June 21, 2012.

The data set is from federal district court decisions interpreting or applying the USA PATRIOT ACT. Using WESTLAW and a time frame from September 2001 through January 2011, a master list of 301 citations was generated by using a generic search term in the DCT (U.S. District Courts) sub-file database: “107-56” “107 PL 56” “107 P.L. 56” “115 STAT 272” “USA PATRIOT ACT” “PATRIOT ACT” “UNITING AND STRENGTHENING AMERICA BY PROVIDING APPROPRIATE TOOLS REQUIRED TO INTERCEPT AND OBSTRUCT TERRORISM” & DA(AFT 9/11/2001).

Weinberg, Pedahzur, and Hirsch-Hoefler's (2010:786–87) synthesis of academic writings defining the term “terrorism” produced this “consensus definition” which has “a relatively strong resemblance to the way states, and law enforcement agencies … regard the phenomenon”: It is “a politically motivated tactic involving the threat or use of force or violence in which the pursuit of publicity plays a significant role.” While it is possible to achieve a consensus definition of terrorism among experts in the field of study, it nonetheless suffers from inherent ambiguity.

Poole and Rosenthal's (1997) common space score uses roll-call votes to create an ideological score for every senator and representative, with −1 being the boundary for the most extremely liberal, +1 being the boundary for most extremely conservative. The value assigned to a legislator is common throughout the course of his or her career, which means that the common space score for each legislator remains the same in different congressional sessions.

When both senators are in the president's party, Hendershot and Tecklenburg (2011) use the difference between the president's ideology and the more distant senator. Although this method is reasonable, we elect to stick with Giles, Hettinger, and Pepper's (2001) method of using the mean between the two senators, with the understanding that senatorial courtesy rotates between the two senators.

Giles, Hettinger, and Pepper's (2001) study of judges on the U.S. Circuit Court of Appeals uses the ideology of the state's elite when senatorial courtesy is not in effect. However, state elites are more likely to be involved in appeals court nominations than in district court nominations. Moreover, Johnson and Songer (2002) find no significant relationship between state elite ideology at the time a U.S. District Court judge is appointed and that judge's decisions. Consequently, like Collins, Manning, and Carp (2010) and Hendershot and Tecklenburg (2011), we elected to use the president's score when there was no senatorial courtesy.

One magistrate in our data set, James K. Bredar, was subsequently appointed to the District Court of Maryland by President Obama in 2010. Consequently, he does have a common space score; however, to keep consistent with the predicted ideology of magistrates at the time a case is decided, we still used the mean ideology for the district in the cases that Bredar decided as a magistrate.

When there were vacancies in the Senate, we adjust the median point accordingly.

Previous research has employed the median member of the committee in question instead of the mean of all the committee members like we did (Owens, 2011; Segal, 1997; Spiller and Gely, 1992). However, those studies encompass long time periods, whereas our study contains a much shorter time frame. Therefore, we found little variation on the median committee member of the Senate Intelligence Committee, despite more frequent changes in committee membership. Consequently, it was preferable for us to employ the mean measure, which had more variation. Moreover, despite the fact that the Intelligence Committee variable is not significant when using the mean or the median, the results are stronger when using the mean.

We also considered using a similar variable capturing the mean common space score of the Senate Judiciary Committee, which could influence the behavior of federal district court judges. However, including this variable created a multicollinearity problem. The Pearson's r between the government ideology variable and the Senate Judiciary Committee ideology variable is 0.818 (p < 0.0005), whereas the Pearson's r between the government ideology variable and the Senate Intelligence Committee ideology variable is only 0.091 and not significant (p = 0.347). Moreover, even though SPSS does not compute tolerance or Variance Inflation Factors (VIF) statistics for logistic regression, we tested for multicollinearity by running an OLS regression with the same dependent and independent variables used in the logistic regression (Menard 1995:66). In a model without the Senate Judiciary Committee ideology variable, the tolerance for the government ideology variable is 0.631, which does not indicate multicollinearity. However, when the Senate Judiciary Committee ideology variable is included in the model, the tolerance for the Senate Judiciary Committee ideology variable is 0.159 and the tolerance for the government ideology variable is 0.153. These figures indicate multicollinearity.

Because SPSS does not compute tolerances for logistic regression, we used OLS regression with the same independent and dependent variables to test for multicollinearity (Menard 1995:66). These tolerances were all far above 0.2, which demonstrates that multicollinearity is not a problem for this model.

Because the SPSS Logistic Regression Function does not compute reduction of error, we computed it with the following formula: 100 * (% classified correctly − % in the modal category) / (100 − % in the modal category).

The formula for computing the percentage interpretation is 100 * (odds–ratio – 1; DeMaris, 1992: 46).