CONTINGENT EXERCISE: A MILD BUT POWERFUL PROCEDURE FOR SUPPRESSING INAPPROPRIATE VERBAL AND AGGRESSIVE BEHAVIOR

Two single-subject experiments were conducted in public school classrooms for severely emotionally disturbed children. Both experiments investigated the effects of a treatment requiring a child to exhibit a simple exercise task after a verbal or aggressive response, using reversal and multiple-baseline designs. The independent variable, contingent exercise, required standing up and sitting on the floor five to ten times contingent on an inappropriate behavior. It was found that contingent exercise was easy to carry out, and following it, the child quickly returned to the learning task that had been interrupted by the inappropriate behavior. The contingent exercise procedure required a minimum of prompting or manual guidance. Although contingent exercise was not topographically related to the inappropriate response, it decreased those responses dramatically. The results suggested that contingent exercise was not only more powerful than DRO but also could be administered independently. It was concluded that contingent exercise may constitute an alternative procedure that can be used by therapists confronted with severely abnormal behaviors. It would appear to be particularly relevant in settings where procedures such as timeout and painful consequences find restricted use. Finally, a number of guidelines have been proposed as a safeguard against the misuse of this mild but powerful procedure.     

Levels in the identifying of orientation by preschool children

A new approach to understanding the perception of orientation proposes that the first level in the processing of orientation information results in perceiving whether a shape is upright or nonupright; nonupright orientations are not distinguished from each other. As predicted, children of 3 and 4 years discriminated upright from nonupright pictures more readily than they discriminated the nonuprights (upside down and sideways) from each other. The angular differences between the pairs of discriminanda (whether 90 or 180°) had no effect on performance. the difficulty of distinguishing between nonuprights could not be attributed simply to a lack of attention to the dimension of orientation. The theory appears to have considerable generality, and provides a basis for understanding (1) perception of the orientation of both realistic and geometric shapes, and (2) similarities in orientation perception observed between children and adults.     

Two classes of models for magnitude estimation

One class of models assumes that presentation of a signal results in an internal representation as a random variable. Depending on whether the signal is close to or far from the preceding signal, the variance of the representation is smaller or larger. Responses are determined largely by this random variable; however, when the signal is close to the preceding one, the response is generated by modifying the representation multiplicatively by some function of the ratio of the previous response to its representation. Power and linear functions are explored. The form of the random variable is assumed to be that arising from either the timing or the counting model operating on a Poisson process. Detailed analyses are carried out successfully only for the timing model with neural sample sizes independent of intensity; however, the data require the sample to increase with intensity. The linear response function coupled with the constant sample size counting model appears somewhat viable, but detailed calculations are very difficult to carry out. The second class of models postulates a power function relation between magnitude estimates and signals intensity for which the exponent is a Gaussian distributed random variable and the unit is the product of two log normal random variables. Again we assume an attention band such that succesive stimuli that are widely separated in intensity lead to independent samples of the random variables while a variety of assumptions is explored for successive stimuli that are near each other in intensity. Although they each give rise to the qualitative features of the data, estimates of parameters are sufficiently inconsistent that we are led to reject all of the submodels studied.     

Variability and sequential effects in cross-modality matching of area and loudness

Individual subjects' performance was examined for cross-modality matching (CMM) of loudness to visual area, as well as for magnitude estimation (ME) of the component continua. Average exponents of power functions relating response magnitude to stimulus intensity were .73 for area, .20 for loudness, and 2.44 for CMM. Predictions of the CMM exponent based on ME were higher than the empirical values, whereas more accurate predictions were made from magnitude production exponents obtained in a previous study. Sequential dependencies were assessed by comparing the response on trial n to the response on trial n--1. The coefficient of variation of the response ratio Rn/Rn-1 was systematically related to the stimulus ratio Sn/Sn-1 for both area and loudness. The coefficient was lowest for ratios near 1 and increased for larger or smaller values. For CMM, the coefficient of variation appeared to be independent of stimulus ratios. The correlation between log Rn and log Rn-1 was also related to Sn/Sn-1 for both ME and CMM. The correlation was highest when Sn/Sn-1 was 1 and dropped to 0 with increasing stimulus separation, but CMM yielded a shallower function than ME.     

Variability of magnitude estimates: A timing theory analysis

Three procedures for magnitude estimation were investigated, and a sufficient number of responses were obtained to make reasonable estimates of both the mean and variance of the responses. The conventional magnitude estimate procedure, without a standard signal, appeared to produce the most sensible data. The best method of establishing the central tendency of the data appears to be the plot of the mean ratio of successive responses against the intensity ratio of the corresponding signal intensities. When this is done, the average response ratio increases roughly as a power function of the signal ratios. The coefficient of variation, σ/m, varies from about 0.1 for small signal ratios and increases to 0.3 at about 20 dB and greater signal separations. The distribution of response ratios appears to be reasonably well approximated by a beta distribution. The change in σ/m with signal ratio is suggestive of an attention mechanism in which the sample size depends on the location of the attention band. The ratio estimation procedure suffers badly from discrete number tendencies.     

Detection of auditory signals presented at random times: III

Reaction times to a pure tone in noise were measured. Throughout, the time from the warning signal to the reaction signal was exponentially distributed, and the signal was response terminated. Response criterion, signal intensity, and mean foreperiod wait were varied. A model that assumes a Poisson sensory transduction, a pulse-activated decision process, and an additive bounded residual process was tested. It was concluded that the assumed decision process was in error. Among the empirical results, the dependence of mean reaction time on signal waits was shown to depend largely on the average wait, not the actual one, and that this relationship between mean reaction time and average stimulus wait increased for strong signals and decreased for weak ones.     

Classification of concatenation measurement structures according to scale type

A relational structure is said to be of scale type (M,N) iff M is the largest degree of homogeneity and N the least degree of uniqueness (Narens, 1981a, Narens, 1981b) of its automorphism group.Roberts (in Proceedings of the first Hoboken Symposium on graph theory, New York: Wiley, 1984; in Proceedings of the fifth international conference on graph theory and its applications, New York: Wiley, 1984) has shown that such a structure on the reals is either ordinal or M is less than the order of at least one defining relation (Theorem 1.2). A scheme for characterizing N is outlined in Theorem 1.3. The remainder of the paper studies the scale type of concatenation structures 〈X, ?, ° 〉, where ? is a total ordering and ° is a monotonic operation. Section 2 establishes that for concatenation structures with M>0 and N<∞ the only scale types are (1,1), (1,2), and (2,2), and the structures for the last two are always idempotent. Section 3 is concerned with such structures on the real numbers (i.e., candidates for representations), and it uses general results of Narens for real relational structures of scale type (M, M) (Theorem 3.1) and of Alper (Journal of Mathematical Psychology, 1985, 29, 73–81) for scale type (1, 2) (Theorem 3.2). For M>0, concatenation structures are all isomorphic to numerical ones for which the operation can be written $\text{x°y = yf(}\text{x}\text{y}\text{)}$, where f is strictly increasing and $\text{f(x)}\text{x}$ is strictly decreasing (unit structures). The equation f(x^?)=f(x)^? is satisfied for all x as follows: for and only for ? = 1 in the (1,1) case; for and only for ?=kⁿ, k > 0 fixed, and n ranging over the integers, in the (1, 2) case; and for all ?>0 in the (2, 2) case (Theorems 3.9, 3.12, and 3.13). Section 4 examines relations between concatenation catenation and conjoint structures, including the operation induced on one component by the ordering of a conjoint structure and the concept of an operation on one component being distributive in a conjoint structure. The results, which are mainly of interest in proving other results, are mostly formulated in terms of the set of right translations of the induced operation. In Section 5 we consider the existence of representations of concatenation structures. The case of positive ones was dealt with earlier (Narens & Luce (Journal of Pure & Applied Algebra27, 1983, 197–233). For idempotent ones, closure, density, solvability, and Archimedean are shown to be sufficient (Theorem 5.1). The rest of the section is concerned with incomplete results having to do with the representation of cases with M>0. A variety of special conditions, many suggested by the conjoint equivalent of a concatenation structure, are studied in Section 6. The major result (Theorem 6.4) is that most of these concepts are equivalent to bisymmetry for idempotent structures that are closed, dense, solvable, and Dedekind complete. This result is important in Section 7, which is devoted to a general theory of scale type (2, 2) for the utility of gambles. The representation is a generalization of the usual SEU model which embodies a distinctly bounded form of rationality; by the results of Section 6 it reduces to the fully rational SEU model when rationality is extended beyond the simplest equivalences. Theorem 7.3 establishes that under plausible smoothness conditions, the ratio scale case does not introduce anything different from the (2, 2) case. It is shown that this theory is closely related to, but somewhat more general, than Kahneman and Tversky's (Econometrica47, 1979, 263–291) prospect theory.     

Validation of the Working Alliance Inventory–Observer Short Version with male intimate partner violence offenders

The working alliance is a key element to increase intimate partner violence (IPV) offenders’ motivation, adherence to treatment, and active participation in batterer intervention programs (BIPs). The objective of the present study is to assess the psychometric properties and factor structure of the Working Alliance Inventory–Observer Short Version (WAI-O-S) with a sample of IPV offenders. The sample was 140 men convicted for IPV and court-mandated to a community-based BIP. Inter-rater agreement and reliability were evaluated by computing the intraclass correlation coefficient. To test the latent structure a Bayesian confirmatory factor analysis approach was used. To test criterion-related validity, the WAI-O-S factorial scores were correlated to protherapeutic behavior, stage of change and motivation to change. The WAI-O-S showed an adequate reliability. Results from Bayesian confirmatory factor analyses showed two first-order factors (Bond and Agreement), and a second-order factor (General working alliance) explaining the relationship between the first-order factors. Results also support the validity of this instrument. The availability of reliable and valid observational measure of the working alliance provides a useful tool to overcome self-report measurement limitations such as social desirability, deception, and denial among IPV offenders.