Background
This paper discusses the development and validation of the Scale of Perceived Power Imbalance (SPPI), an instrument designed to measure children’s experience of power imbalance associated with bullying. The definition of bullying provides a basis for the development of the measurement tool [
1]. For the purpose of this study, school bullying is defined as a form of aggression that is distinguished by
repeated physical or emotional harm within a relationship of
power imbalance [
2]. This differs to the definition provided by Olweus [
3], who included
intent to harm in the definition of bullying. The criteria of intent differentiates purposeful acts of aggression from accidental harm [
1]. It has been proposed that intentionality is understood within the context and goals of bullying: that harm is intended by the perpetrator, is perpetrated within the social dynamic of the peer group, and is perceived by the victim of bullying [
4]. Based on a legal framework, it has been proposed that the judgement of intent rests on the likelihood that a reasonable person would foresee that aggressive behavior would result in harm [
1,
5]. This criteria is difficult to apply as intent is not easily observed [
1,
6]. For these combined reasons, the criteria of intent is not included in the definition of bullying for this paper.
The concept of repetition as an essential criteria of bullying has also been questioned by researchers, because single acts of aggression can provide an ongoing threat resulting in long term physical or emotional harm [
5,
7]. However, the uniform definition of school bullying as unwanted aggression that is repeated and involves a power imbalance is accepted internationally [
8,
9]. Consistent with this, repetition of aggression has been empirically associated with an experience of significantly greater threat and harm by targeted children than those who experienced aggression without repetition [
10‐
12]. Children who are bullied perceive that they do not have the power to stop repeated aggression, contributing to an increase in harm [
12]. The repeated acts of aggression that occur in a power imbalanced relationship place a load on neuroendocrine pathways that respond to stress, increasing the risk of poor health, learning and developmental outcomes [
13]. Outcomes associated with bully victimization include fear, loss of hope, anxiety, depression and suicidal ideation [
14]. The core concept that differentiates bullying from aggression is, therefore, the abuse of power by the perpetrator, and the experience of a power imbalance by the victim [
4]. Despite the importance of the concept of power imbalance in bullying research, it is reported in recent literature that power imbalance has not been measured effectively, resulting in the inaccurate reporting of aggression as bullying [
15].
Reports of bullying begin to increase in grade 4 to 6 (age 8 to 12) as children develop the cognitive capacity for self-reflection and place increasing value on social hierarchies [
16,
17]. The concept of power imbalance as central to the definition of bullying has been driven by researchers, and children themselves may not consider repetition or power imbalance when talking about bullying [
18]. They are, therefore, likely to report all acts of aggression as bullying [
19]. With this in mind, researchers have sought to include the constructs of repetition and power imbalance in their self-report bullying instruments. Some researchers include a definition and ask children to consider the definition while answering questions [
20]. There is evidence that many children fail to apply the definition of bullying correctly when answering the questions that follow and are prone to report aggressive behavior as bullying [
15]. Other children may avoid answering questions truthfully when they read the word
bully within the definition because of the stigma and shame associated with being a victim of bullying [
21]. In this way the definition-based approach is contrary to conventional self-report measurement in psychology where questions ask about very specific behaviors and experience to discourage bias associated with respondents giving socially-desirable answers [
22].
A second method of assessment seeks to increase the accuracy of measurement by asking children who report frequent victimization direct questions to assess perceived power imbalance. This is referred to as the behavioral-based method of assessment [
23]. For example, Hunter et al. [
10] aimed to differentiate bullying from victimization using three individual items to determine if the aggressor was physically stronger, in a bigger group, or more popular than the child completing the survey (ages 8–13). Concurrent and discriminant validity of the instrument were supported; students who experienced power imbalance perceived more threat and loss of control, and were at greater risk of depression. Similarly, the Californian Bully Victim Scale (CBVS) included three items to measure power imbalance from the perspective of the victim, asking “how popular, smart in schoolwork, and physically strong” the aggressor was [
23]. Predictive validity was established; students who experienced power imbalance reported lower connectedness to school, life satisfaction, and hope. Five items were added to the CBVS: “how likeable, good looking, athletic, old, and how much money” the aggressor had in comparison to the victim [
24]. These authors assessed the validity of a definition versus behavioral approach (grade 5 to 9) by comparing responses to one definition-based item from the Olweus Bullying Questionnaire (OBQ) [
3]. Power imbalance was not significantly associated between the two measures. The authors concluded that the definition-based method might not detect some forms of power imbalance [
24]. Notwithstanding, the accuracy of each item to detect the power differential using the behavioral approach remains unclear [
25]. Cornell and Limber [
26] claim that a satisfactory method to measure the power differential is yet to be identified.
Volk et al. [
4] recommended that new instruments intended to measure bullying are based on insights gained from qualitative studies, and that specific forms of power are presented and validated. We have designed an instrument that uses this approach to measure the individual perception of power imbalance associated with repeated victimization at the level of the dyad. Our approach was innovative in that we worked with children aged 8 to 12 years to design an instrument and used factor analysis to explore the psychometric fit of items designed to measure the power imbalance component of bullying. The new instrument was implemented in an online survey.
The aim of this research was to establish the reliability and validity of the new instrument, named the Scale of Perceived Power Imbalance (SPPI). On the basis of previous work [
27] we anticipated a negative association between perceived peer support and perceived experience of power imbalance by children who reported frequent victimization. We expected to find a moderate correlation of the SPPI with an existing measure of bully victimization, and that the SPPI would demonstrate reliability over time and invariance by gender, grade, and school.
Results
Exploratory factor analysis
The following number of children reported frequent victimization by each scale: APRI verbal (
n = 48), APRI physical (
n = 28), APRI social (
n = 36), and PECK verbal-relational (
n = 55). When missing data were deleted listwise, the number of participants who answered the power imbalance scale in response to the APRI verbal (
n = 44)
, physical (
n = 28), and social scales (
n = 34) was insufficient to run EFA, based on a minimum of five cases per item [
41] or the minimum Kaiser criterion for sampling adequacy of .60 [
42]. Results of the EFA are therefore reported on the power imbalance items that followed the PECK Verbal-relational scale (
n = 51). Data were normally distributed. EFA in SPSS resulted in a KMO of .576 suggesting a “mediocre” fit of the data [
55] and resulted in three eigenvalues ≥ 1. Because the eigenvalue criterion can lead to the extraction of too many factors [
48] parallel analysis was calculated, resulting in a 2-factor structure. EFA of the 2-factor model in MPlus resulted in acceptable fit (
normed χ2 = 1.25, RMSEA = .070 [90%CI = .00 to 0.148], CF1 = .942, SRMR = .119). The 90% confidence interval was wide; it included the value of .05 suggesting that the result reflected a small sample size. Factor loadings are shown in Table
2, the correlation between factors was −.039. Item 5 “trying to be more popular” did not load onto either factor.
Table 2
EFA of the power imbalance items in MPlus
Item 1 |
0.491
| −0.285 | Really smart? |
Item 2 |
0.905
| −0.125 | Good looking? |
Item 3 | 0.027 |
0.556
| Older than you? |
Item 4 |
0.513
| 0.193 | Good at sport? |
Item 5 | 0.200 | 0.139 | Trying to be more popular? |
Item 6 |
0.764
| 0.002 | In the most popular group? |
Item 7 | 0.013 |
0.762
| Much stronger than you? |
Item 8 | −0.156 |
0.902
| Bigger than you? |
Item 9 |
0.813
| 0.271 | With a group of students? |
Confirmatory factor analysis
CFA of the hypothesized 9-item 2-factor model of the SPPI did not provide a good fit to the data. On examination, the item “good at sport” had a low communality when answered in response to either physical/verbal or social victimization (R2 = .183, and .034 respectively), and did not load onto the factor when answered in response to the APRI social victim scale (standardized loading = .184). This was consistent with a high tolerance of the item (.919) indicating the uniqueness of the item. “Good at sport” was excluded from further analysis and an 8-item model of the SPPI was tested for fit.
Baseline model fit of the 8-item SPPI in response to
verbal/physical victimization was acceptable (
n = 146: normed
χ2 =1.96, RMSEA = .081 [90%CI = .041 to .119], CF1 = .906). Standardized factor loadings ranged from .353 “really clever” to .941 “tougher than you”. The
R2 for “really clever”, and “good looking” were low (see Table
3). In response to
social victimization factor loadings of the baseline model of the 8-item SPPI were adequate (
n = 127: normed
χ2 =1.2, RMSEA = .04 [90% CI = .00 to .091], CF1 = .993). The communality of Items “really clever”, and “good looking” remained low (see Table
3). This is addressed in the discussion.
Table 3
CFA of the factor structures of power imbalance items in MPlus
Social power |
Really clever? |
0.353
| 0.124 |
0.419
| 0.175 |
Good looking? |
0.388
| 0.15 |
0.473
| 0.224 |
In the most popular group? |
0.894
| 0.799 |
0.790
| 0.625 |
With a group of students? |
0.624
| 0.389 |
0.680
| 0.462 |
Composite Reliability |
0.668
| |
0.689
| |
Physical power |
Older than you? |
0.475
| 0.226 |
0.646
| 0.418 |
Much stronger than you? |
0.792
| 0.627 |
0.891
| 0.793 |
Bigger than you? |
0.519
| 0.27 |
0.877
| 0.769 |
Tougher than you? |
0.941
| 0.885 |
0.974
| 0.949 |
Composite Reliability |
0.789
| |
0.915
| |
Multi-group invariance
Multi-group analysis of the 8-item SPPI was conducted with the MLR estimator using the baseline models of best fit to assess invariance of the instrument. In each data set, the configural model of the SPPI answered in response to children’s self-report of verbal/physical victimization was inadmissible because of a negative residual variance for the item “in the most popular group” (gender,-.257; grade at school, −.043). The unstandardized factor loading for gender was high for boy (39.522, standardized =1.451), and for Grade 5–6 (9.647, standardized =1.085).
Invariance was therefore assessed on the 8-item SPPI that was answered in response to self-report of social victimization. Invariance of the mean and covariance structures was demonstrated across gender as evidenced by a non-significant corrected
χ2 between the nested model and the comparison model at each specified level of the model (
p > .05). Invariance of the mean and covariance structures was likewise demonstrated between grade at school (Grade 4, and Grades 5–6) as evidenced by a non-significant corrected
χ2 between the nested model and the comparison model at each specified level of the model (
p > .05) (reported in Table
4).
Table 4
Multigroup analysis of the SPPI in response to social victimization
Power Social | Total group analysis (n = 127) | Factor1 by items 1, 2, 6, and 9; Factor 2 by items 3, 7, 8, and 10. | |
CFA (n = 127) | 23.152 | 19 | .975 | 0.056 | 25 | .041 [.00–.092] | | | | | | | |
Invariance of covariance structures - gender |
CFA girl (n = 72) | 23.659 | 19 | .949 | 0.074 | 25 | .058 [.00–.125] | | | | | | | |
CFA boy (n = 54) | 22.929 | 19 | .945 | 0.076 | 25 | .062 [.00–.140] | | | | | | | |
MGGender-M1 | 46. 593 | 38 | .947 | 0.075 | 50 | .060 [.00–.112] | Configural | 1.0608 | | | | | |
-M2 | 53.707 | 44 | .941 | 0.077 | 44 | .059 [.00–.108] | Factor loadings | 1.0464 | 2 subtract 1 | .9552 | 7.0740 | 6 |
>.20
|
-M3 | 54.086 | 47 | .957 | 0.078 | 41 | .049 [.00–.100] | Structural model | 1.0430 | 3 subtract 2 | .9931 | .2142 | 3 |
>.20
|
Invariance of mean and covariance structures |
-M4 | 54.149 | 46 | .950 | 0.079 | 42 | .053 [.00–.103] | Factor loadings | 1.0412 | 4 subtract 1 | .9481 | 7.5848 | 8 |
>.20
|
-M5 | 61.099 | 54 | .957 | 0.087 | 34 | .046 [.00–.095] | Intercepts | 1.0338 | 5 subtract 4 | .9913 | 6.8441 | 8 |
>.20
|
-M6 | 59.467 | 50 | .942 | 0.081 | 38 | .055 [.00–.102] | Latent means | 1.0269 | 6 subtract 4 | .8625 | 4.8365 | 4 |
>.20
|
Invariance of covariance structures - grade |
CFA grade 4 (n = 50) | 14.694 | 19 | 1.000 | 0.067 | 25 | .000 [.00–.091] | | | | | | | |
CFA grade 5_6 (n = 77) | 21.438 | 19 | .977 | 0.063 | 25 | .041 [.00–.111] | | | | | | | |
MGGrade-M1 | 36.274 | 38 | 1.000 | 0.065 | 50 | .000 [.00–.082] | Configural | 1.0653 | | | | | |
-M2 | 38.791 | 44 | 1.000 | 0.070 | 44 | .000 [.00–.068] | Factor loadings | 1.0497 | 2 subtract 1 | .9509 | 2.1834 | 6 |
>.20
|
-M3 | 40.222 | 47 | 1.000 | 0.073 | 41 | .000 [.00–.061] | Structural model | 1.0312 | 3 subtract 2 | .7599 | .9976 | 3 |
>.20
|
Invariance of mean and covariance structures |
-M4 | 39.735 | 46 | 1.000 | 0.071 | 42 | .000 [.00–.063] | Factor loadings | 1.0288 | 4 subtract 1 | .8554 | 2.6147 | 8 |
>.20
|
-M5 | 47.850 | 54 | 1.000 | 0.083 | 34 | .000 [.00–.062] | Intercepts | 1.0213 | 5 subtract 4 | .9782 | 8.1681 | 8 |
>.20
|
-M6 | 44.132 | 50 | 1.000 | 0.077 | 38 | .000 [.00–.064] | Latent means | 1.0400 | 6 subtract 4 | 1.1688 | 4.2932 | 4 |
>.20
|
Test-retest reliability
Test-retest reliability (2-week interval) of the combined ranked score (n = 50) from the APRI and SPPI was indicated by a moderately strong correlation when answered in response to self-report of verbal/physical victimization rs =.773 (p < .001), and a strong correlation in response to social victimization rs =.841 (p < .001).
Validity
Construct validity of the SPPI was assessed in relation to the preceding APRI social victim scale. CFA of the 3-factor model of social victimization, social power, and physical power resulted in a good fit (n = 124: normed χ2 =1.09, RMSEA = .027 [90%CI = .00 to .057], CF1 = .983). Correlations between factors were low suggesting discriminant validity of the two latent factors of power imbalance (.294) and between the APRI social victim scale and power imbalance (social power = .325, physical power = .357).
Discriminant validity was supported by negative correlations between children’s perception of peer support (PPSS; Ladd et al., 1996) and each latent factor of the SPPI: social victimization (physical power = −.292, social power = −.175), CFA of the 3-factor model (n = 120: normed χ2 =1.0, RMSEA = .009 [90%CI = .00 to .041], CF1 = .997); and verbal/physical victimization (physical power = −.138, social power = .035), CFA (n = 140: normed χ2 =1.09, RMSEA = .025 [90%CI = .00 to .046], CF1 = .957).
Convergent validity was supported by a moderate correlation between the screening item of the OBQ [
3] and the two latent factors of the eight-item SPPI (
overt and
social victimization) (
n = 331,
rs =.533). Fifty five per cent of children reported victimization in a relationship of power imbalance by either the SPPI or the OBQ. The overlap in self-report of power imbalance between instruments was 20.2% (see Table
5). Fewer children reported being bullied by the OBQ screening item (23.8%) in comparison to those who reported frequent victimization by the combined ranked score of the 19-item APRI and the eight-item SPPI (50.8%) (see Table
5).
Table 5
Comparison of children’s self-report of victimization with power imbalance by instrument
Did not report victimization with perceived power imbalance by either scale | 149 | 45.0 | 45.0 |
Reported victimization with perceived power imbalance by combined rank score of APRI+SPPI only | 103 | 31.1 | 76.1 |
Reported bully victimization by OBQ only | 12 | 3.6 | 79.8 |
Reported victimization with power imbalance by APRI & SPPI and OBQ | 67 | 20.2 | 100.0 |
An additional finding was that 11% of children who reported frequent social victimization and 8% of those who reported frequent overt victimization did not report an experience of power imbalance by the 8-item SPPI (see Table
6).
Table 6
Comparison of children’s self-report of frequent victimization with and without power imbalance
Overt victimization (n = 335) | 184 | 151 | 141 | 10 | 6.6% |
Social victimization (n = 335) | 202 | 133 | 119 | 14 | 10.5% |
Discussion
Analyses of the items in the SPPI resulted in a two-factor structure. Social power measured the peer-valued characteristics of clever, appearance, athleticism, belonging to the popular group, and being with a group. Physical power measured the physical characteristics of age, size, strength, and toughness. Physical measures of power imbalance are more obvious, thus easier to quantify, than peer-valued characteristics, as evidenced by consistent fit to the data. Likewise, the group aspects of social power displayed consistently high factor loadings in response to children’s report of overt and physical victimization. The peer-valued characteristics of social power, “good at sport,” “clever,” and “good looking,” however, displayed inconsistent factor loadings and low communalities.
The factor
social power measured characteristics that are valued by peers and associated with peer acceptance or belonging. The a priori factor structure was supported by the extant literature. The item “good at sport” did not load onto the social power factor in CFA. This is in contrast to a high response rate of
yes for children answering this item in the SPPI (30% to
verbal/physical victimization and 22% to
social victimization). When Green et al. [
24] previously measured power imbalance by individual items, children also responded most frequently to the item “more athletic” (39.3%). The reliability of the measure of power imbalance was however, not reported by the authors. The low factor loading for “good at sport” in our research does not mean that athleticism is not associated with power imbalance, the low communality and high tolerance (0.919) of the item suggested, however, that it was not related to the other items of social power [
51].
Felix et al. [
23] found that the item “smart in school work” was possibly not a good question to address power imbalance based on a small response rate to the item. Despite this the item was retained in the CBVS [
24]. Similarly, in our research less than 20% of the variance of the item “really clever” was explained by the
social power factor to which it was linked. Items such as “good at sport” and “clever”, and their opposites, might reflect an academic or ability-related form of power imbalance. This warrants further investigation. Moreover, in focus groups, children referred to smart as “getting their way out of trouble” by hiding the behavior from adults [
30]. This social manipulation resulted in feelings of hopelessness and the inability to escape the situation by the victim, increasing the power of the perpetrator over the victim. This is consistent with recent qualitative research. Teachers may not recognize perpetration of aggression by popular students and may even place the responsibility for victimization on the targeted student [
56]. Resilience is fostered when children receive social support. It is, however, possible that some children perceive that the teacher is supporting the child who is doing the bullying and is therefore not available as a source of social support [
6]. This form of power is difficult to assess, and has been difficult to quantify, but must not be ignored because it is associated with poor health outcomes [
14]. For this reason, we propose that further investigation into “smart” or “clever” as a form of power imbalance might inform the development of prevention strategies, including the promotion of resilience.
The item “good looking” was used as a measure of power in the revised CBVS [
24] and was found to be associated with power in focus group analysis [
30]. However, we found that only 22% of the variance of the item was explained by the
social power factor to which it was linked. In focus groups, children referred to “looks”, but beyond appearance,
looks also related to clothes, shoes, smart phones, and possessions. These are all consistent with the attribution of appearance to social power [
57]. These items were kept because: a) they may be providing important information regarding the latent construct; and b) the amount of variance they contribute may be important, even if their factor loading was consistently low [
58].
Cascardi et al. [
2] have suggested that popularity is potentially a superficial feature of power imbalance. Our initial invariance analysis models were not identified for the SPPI items answered in response to the
verbal/ physical victimization sub-scale of the APRI due to very high factor loadings on the item “in the most popular group”. These results suggest that boys were likely to experience a very high experience of power imbalance when the aggressor belonged to the popular group. Girls in Grades 5 and 6 were similarly more likely to experience a high power imbalance if the aggressor belonged to the most popular group. This is consistent with qualitative findings that popularity is a key influence on bullying within the group [
59].
Preliminary support was found for the construct validity of the SPPI. Discriminant validity was supported by the strong 3-factor structure of the PPSS, SPPI
physical power sub-scale, and
social power sub-scale. As expected the correlation between perceived peer support and perceived power imbalance was very low, suggesting that children who felt supported by peers experienced lower rates of victimization with power imbalance. Consistent with previous research we found that double the number of children reported being bullied when they answered questions about individual types of bullying compared to the screening item of the OBQ [
39].
Limitations
This study used a single method of anonymous self-report and there was no verification of victimization by a different source, for example peers or teachers. There are, however, ethical considerations with peer nomination [
20]. Agreement between the multiple approaches of quantitative analysis did support the 2-factor solution of power related to peer-valued characteristics and physical characteristics. However, thematic analysis of focus group discussion resulted in three forms of power imbalance reflecting physical characteristics (age), peer valued characteristics, and group membership and position [
30]. Future research might benefit from exploring group and peer-valued characteristics of power as different constructs.
The SPPI was developed for the local context, however the qualitative phase of the study was reported in the context of prior international research [
30]. There is a great need to reduce the harm of bullying in schools [
60]. Victims of bullying experience a power imbalance that hinders their perceived ability to stop the repeated aggression. For this reason it is necessary to continue investigating the nature and measurement of power imbalance.
A limitation of this study is the focus on power imbalance associated with traditional forms of bullying and not cyberbullying. The form of power imbalance is likely to differ between traditional and cyberbullying [
1]. Bauman et al. proposed that the imbalance of power associated with cyberbullying should not be assessed by self-report where possible because of the difficulty in inferring power imbalance from a subjective response to online communication [
1]. In contrast, it is widely recommended that self-report will provide the most accurate measure of the power imbalance in traditional bullying [
9,
20,
38]. This research focused on providing evidence on the reliability and validity of a specific measurement technique to attain children’s self-reported experience of power imbalance associated with traditional forms of bullying [
25].
Conclusion
There is reasonable agreement among researchers concerning the influence of physical factors, such as age, size, and toughness, on power imbalance. These have long been acknowledged in the literature and were found to have a strong factor structure in the analysis undertaken in this study. The influence of social factors, such as being good looking, smart, or popular, on power imbalance is much less clear for a number of reasons. Firstly, social forms of power are subtle in nature and not easily recognized by authority figures. Secondly, social power draws strength through peer group dynamics. This is supported in this study by the strong factor loadings of items that were specific to the peer-group. Thirdly, the factors that comprise social power are often highly valued by peers. The characteristics, themselves, are neutral; it is the status that peers attribute to these characteristics that confers power on those who possess them. Further research is required to better understand the influence of social factors on power imbalance. Specifically, we need to understand how social power reflects: 1) the dominance goals of the perpetrator; 2) group functioning; and 3) the perceived inability of the targeted child to overcome repeated victimization.
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