Bimanual response grouping in dual-task paradigms

Abstract

In three experiments we measured response time (RT) and peak force (PF) to investigate the grouping of left- and right-hand key press responses in a dual-task paradigm involving two independent go/no-go tasks. Within each task, a go stimulus within one of two modalities (i.e., visual versus auditory) required a response by one hand. In Experiment 1 with simultaneous go stimuli in the two tasks, responses appeared to be grouped in approximately 75–80% of trials, compared with nearly 100% grouping in a single-task condition requiring bimanual responses to the onset of any stimulus in either modality. In Experiment 2 with stimulus onset asynchronies (SOAs) of 0–400 ms between the two go stimuli, response grouping clearly declined as SOA increased, although some grouping was still evident even at the longest SOA. The same pattern was observed in Experiment 3 with the same range of SOAs but unpredictable stimulus order, suggesting that grouping is not strongly dependent on prior knowledge of the likely response order. These results emphasize the pervasiveness of response grouping in bimanual dual-task RT paradigms and provide useful clues as to its nature.

The research was supported by an Otago Research Grant to Jeff Miller. The authors wish to thank two anonymous reviewers for helpful comments on an earlier version of the manuscript.

Notes

¹ Davis (1959) used a different approach, testing for response grouping by examining the correlation of RT ₁ with RT ₂ across trials. If participants group responses, then there should be a strong positive correlation; in the limit of simultaneous responses, in fact, the correlation should be perfect. Davis found a correlation of only .77 at an SOA of 50 ms and concluded that this was not high enough to suggest response grouping. More detailed consideration suggests that the RT ₁/RT ₂ correlation is not a very clear-cut diagnostic of response grouping, however (cf. Navon & Miller, 2002; Ulrich & Miller, 2007). For one thing, this correlation need not be extremely strong if responses are only grouped on some small proportion of trials, say 20–50%. Indeed, within some models, grouping responses on a proportion of trials can even decrease the RT ₁/RT ₂ correlation relative to what would be expected with no grouping (Ulrich & Miller, 2007). Moreover, even a small proportion of grouped trials might significantly affect mean RTs, regardless of the resulting RT ₁/RT ₂ correlation. Finally, even a rather high RT ₁/RT ₂ correlation would not necessarily indicate that responses were ever grouped, because there are a variety of other mechanisms that could induce positive RT ₁/RT ₂ correlations (e.g., Navon & Miller, 2002; Pashler, 1994a, 1994b).

² The race model inequality (Miller, 1982) was also tested to see whether simple RT responses to redundant bimodal stimuli were faster than could be explained in terms of a race model. Using the method described by Ulrich, Miller, and Schröter (in press), tests were carried out for the 5, 15, 25, . . ., 95 percentile points of the cumulative distribution functions (CDFs) for the single- versus redundant-stimulus RT distributions, and the race model was significantly violated (p < .05) in the comparisons for all of the percentiles in the range of 15–65.

³ We also analysed response force–time functions in several other ways to look for force measures that would be useful in discriminating between grouped and ungrouped responses, but we had no success at that. For example, several of these analyses compared the full force–time profiles of the two bimanual responses (e.g., by correlating their values across time points), based on the intuition that grouped responses should have more similar profiles than ungrouped responses. In all such analyses, the force-based measures produced only slightly more evidence of grouping in the simple RT blocks than in the dual-task blocks, with largely overlapping distributions like those shown for the IPF measure in Figure 1.

⁴ A total of 2 participants were excluded from the computation of means with long IRTs, and 1 was excluded from the computation of means with short IRTs, because these participants had no trials with appropriate IRTs in at least one of the conditions shown in the figure.

⁵ Because the SOAs were symmetric around zero, the distinction between Tasks 1 and 2 is arbitrary for single-stimulus trials and for trials with SOA = 0. Therefore, for these conditions identical Task 1 and Task 2 mean RTs were computed averaging across both the visual and auditory stimuli.

⁶ A total of 4 participants were excluded from the computation of means with long IRTs, and 4 others were excluded from the computation of means with short IRTs, because they had no trials with appropriate IRTs in one or more conditions.