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On the Diagnosticity of Multiple-Witness Identifications

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Law and Human Behavior

Abstract

It is not uncommon for there to be multiple eyewitnesses to a crime, each of whom is later shown a lineup. How is the probative value, or diagnosticity, of such multiple-witness identifications to be evaluated? Previous treatments have focused on the diagnosticity of a single eyewitness’s response to a lineup (Wells and Lindsay, Psychol. Bull. 3 (1980) 776); however, the results of eyewitness identification experiments indicate that the responses of multiple independent witnesses may often be inconsistent. The present paper calculates response diagnosticity for multiple witnesses and shows how diagnostic probabilities change across various combinations of consistent and inconsistent witness responses. Multiple-witness diagnosticity is examined across variation in the conditions of observation, lineup composition, and lineup presentation. In general, the diagnostic probabilities of guilt were shown to increase with the addition of suspect identifications and decrease with the addition of nonidentifications. Foil identification results were more complicated-diagnostic of innocence in many cases, but nondiagnostic or diagnostic of innocence in biased lineups. These analyses illustrate the importance of securing clear records of all witness responses, rather than myopically focusing on the witness who identified the suspect while ignoring those witnesses who did not.

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Notes

  1. There is a potentially important distinction between cases and suspects. Yuille and Tollestrup (1992) and Tollestrup et al. (1994) reported results using cases as the unit of analysis, whereas Wright and McDaid (1996) used the suspect as the unit of analysis. Each case may involve multiple perpetrators and hence multiple suspects. The calculation of the Yuille–Tollestrup result is based on 622 cases rather than 626 shown in their Table 1 because in four homicide cases there were no witnesses. Calculations for Tollestrup et al. (1994) are based only on robbery cases, and one case that involved no witnesses is excluded for a total of 76. Calculations for Wright and McDaid are estimated from their Fig. 1 (p. 77).

  2. The calculation considers response categories rather than individual responses. Two witnesses could select a foil, but two different foils.

  3. The original treatment of diagnosticity (as developed by Wells and Lindsay 1980, and used in Wells and Turtle 1986, and in Wells and Olson 2002) used a ratio, p(R|S = G)/p(R|S ≠ G), representing the likelihood that the response would occur when the suspect was guilty versus innocent. Using the likelihood ratio index, the value of 1.0 indicates no diagnosticity, 2.0 means that the response was twice as likely if the suspect was guilty than if the suspect was innocent, and so on. Here, we have chosen to use the index p(R|S = G)/[p(R|S = G) + p(R|S ≠ G)], which represents diagnosticity as a probability rather than as a ratio. A probability of .50 indicates no diagnosticity, values between .50 and 1.0 indicate degrees of diagnosticity of guilt, and values between 0.0 and .50 indicate degrees of diagnosticity of innocence. The expression of diagnosticity as a probability rather than as a ratio has two advantages. First, the likelihood ratio index required the use of the obverse ratio, p(R|S ≠ G)/p(R|S = G), for responses that were diagnostic of innocence, an unnecessary step when using a probability as the index of diagnosticity. Second, the likelihood ratio is awkward (mathematically) for combining responses from multiple witnesses because it requires the multiplication of ratios (which can expand to be quite large), whereas the use of the probabilistic index of diagnosticity requires the multiplication of probabilities (which are constrained between 0 and 1).

  4. The actual distribution of identifications across lineup members is likely to vary as a function of the lineup composition. The analyses carried out here and elsewhere (Clark et al. 2007) have assumed target-absent lineups that were unbiased in their composition. This assumption arises from the fact that in the relevant studies there was no designated innocent suspect, and thus suspect identification rates in TA lineups were calculated by dividing the overall identification rate by the number of lineup members.

  5. The idea of the double-blind lineup, first proposed by Wells (1988) and emphasized strongly in lineup reform recommendations (Wells et al. 1998), would effectively eliminate the problem of the lineup administrator failing to make a clear record of foil identifications. With a double-blind lineup, the lineup administrator does not know which lineup member is the suspect and which are the foils. Hence, the record made by the double-blind administrator would have to record the identification decision with equal veracity regardless of whether it was an identification of the suspect or a foil.

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Authors and Affiliations

  1. Psychology Department, University of California, Riverside, Riverside, CA, 92521, USA

    Steven E. Clark

  2. Iowa State University, Ames, IA, USA

    Gary L. Wells

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  1. Steven E. Clark
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  2. Gary L. Wells
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Correspondence to Steven E. Clark.

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Clark, S.E., Wells, G.L. On the Diagnosticity of Multiple-Witness Identifications. Law Hum Behav 32, 406–422 (2008). https://doi.org/10.1007/s10979-007-9115-7

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