Being involved in bullying as a victim or perpetrator could have deleterious health consequences. Even though there is some evidence that bullies and victims of bullying have a higher risk for drug use, less is known about bystanders. The aim of this research was to study the association between bullying experience  and substance use. We gathered complete information from a nationally representative sample of 36,687 students  attending 756 schools in Chile. We used a self-reported questionnaire which was developed based on similar instruments used elsewhere. This questionnaire was piloted and presented to an expert panel for approval. We used multilevel multivariate logistic regression analyses, controlling for several variables at the individual  and school levels . This study shows that bullies and bully-victims have a high risk for cigarette, alcohol, and cannabis use than bystanders. This is one of the few studies exploring the association between witnessing bullying and substance use. These findings add new insights to the study of the co-occurrence of bullying and substance use. Other factors, such as higher academic performance, stronger school membership, and better parental monitoring reduced the risk of any substance use, while the experience of domestic violence and the perception of social disorganization in the neighborhood, increased the risk. These findings may help the design of preventive interventions. Aggressive behavior at school is a complex phenomenon, and bullying is probably the most common expression of it. Bullying is a major health problem in schools worldwide  in Norway. This study showed that almost 15% of the students were involved in bullying Olweus, . Since tHowever, as was mentioned earlier, most bullying research has been conducted in developed countries  multilevel modeling is the most appropriate approach to analyze it. However, few studies have explored these associations using multi-level modeling. For instance, Bradshaw et al.  developeMuch less information can be found in Latin American countries , the monthly prevalence in 2015 was found to be 13.5, 1.6, and 0.3%, respectively. Additionally, there has been an increase in the prevalence reported since the previous survey (2012) at least for alcohol and cannabis use , family , and community factors (perception of violence in the neighborhood) at the individual level, and contextual factors  at the school level. To our knowledge, this is the first multilevel study in Latin-America exploring this association, and this is one of the few studies exploring these associations in the different groups that play a role in bullying. Specifically, individuals with no experience of bullying, bystander only, victim only, bully only, and bully-victim. Our general hypothesis is that the exposure to bullying in adolescence, as perpetrator, victims or as bystander, will increase the risk for substance use.This is an analytical cross-sectional study that uses a self-reported survey.Data came from the Forth National Survey of Violence in School Context 2014 (ENVAE 2014) in Chile. This survey used a stratified nationally representative sample of students, attending 7 to 12th grades, in state, subsidized, and private schools in urban municipalities of Chile. This study used simple random sampling, stratified in two stages (schools and students), with a probability proportional to the number of students in each school.The sample frame for this study was 944 schools. It was planned to randomly invite 60 students per school to participate in the survey. All schools were contacted prior to data collection to obtain permission and send consent documents to parents or caregivers. A total of 756 schools agreed to participate, reaching out 45,360 parents and caregivers. Around of 84% of students were given consent and attended the day of the survey. Data collection took place during one school day, on a date pre-determined by the school authorities, between September and December 2014. No replacements were implemented. Finally, complete data from 37,931  students from 754 schools was collected.This secondary analysis study was approved by the Bioethical Committee of Universidad de los Andes (Chile) . It was performed in accordance with the Declaration of Helsinki. All participants provided informed consent.n = 117) was conducted during July 2014 to assess the structure and length, wording, and understanding of the questionnaire. The expert panel approved the final version of the questionnaire.The self-report questionnaire consisted of questions regarding students' demographic features, questions regarding school violence (including bullying experience), academic performance, school norms awareness, school bonding, school membership, perception of peer substance use at school, parental monitoring, family maltreatment, violence and drug-related actions in the neighborhood, and substance use. The expert committee responsible for the study built most of the items and scales. A pilot study , and how many days during the last 12 months prior to the study they had used cannabis . Then, we categorized the answers into two possibilities: 0 = Not user and 1 = User . The 30-day period for tobacco and alcohol use, and the 12-month period for cannabis use are standard and recommended time intervals used in school surveys to define current users (United Nations Office on Drugs and Crime (UNODC), A structured questionnaire covering socio-demographic factors, bullying experiences, other school-related factors, and family and community factors was administered to each student.Socio-demographic features. Information about age, sex, and school year was gathered.Bullying experience. Three independent scales, using a similar set of 9 different situations that students may have been experienced in the current academic year, were used to identify students who had been victims of bullying, bullies, or bystanders (observing bullying actions with no direct involvement). The nine items of the scale for victims were: During the current academic year, how often have other students in your school (a) spread lies or hurtful things about you; (b) hit or kicked you; (c) threatened you; (d) thrown heavy objects at you with the intention of hurting you; (e) ignored or excluded you; (f) insulted you; (g) said mean names to you; and (h) pushed you with the intention of hurting you. Students answered to each event using a 5-point scale . The scales exploring bullying perpetration and witnessing of bullying (bystanders) used the same set of 9 items, but the wording was changed accordingly. We re-arranged these answers and considered a bullying experience if the students responded having experienced any of these actions at least every month. Therefore, three new bullying binary variables were created for each scale: Being a victim ; Being a bully ; and Being a bystander . Later, we combined these three variables into one new variable were students could be identified in mutually exclusive categories: 0 = having no experience of bullying actions , 1 = being a bystander only ; 2 = being a victim only , 3 = being a bully only ; and 4 = being a bully-victim .Self-reported academic performance. In Chile, the grade point average [GPA] scale goes from 1 to 7, where 7 is the highest GPA, possible answers were arranged in six categories from 1 = < 4.5 to 6 = 6.5–7.0). The higher the score, the better the academic achievement.School norms awareness. This is a 6-item Likert scale (1 = strongly disagree to 5 = strongly agree). Items explore several aspects of awareness of school's norms . The higher the score, the higher the awareness of school's norms. This is a unidimensional scale, with internal reliability, alpha = 0.78.School bonding. One item explores this variable asking “generally, how happy are you to go to school?” with 5 potential answers . A higher score represents a better school bonding. This item was taken from a questionnaire used in the World Health Organization cross-national study called “Health Behavior in School-Aged Children” (HBSC) , if the students feel respected by the teachers and/or school staff , and if there are good relationships between students and teachers (“There is trust in the relationship between students and teachers”). Students responded how much they agreed with each statement (1 = strongly disagree to 5 = strongly agree). A higher score means having a higher sense of school membership. This is a unidimensional scale, with internal reliability, alpha = 0.90.Perception of peer substance use at school. This is a 4-item scale exploring the perception of peer substance use at school . For example, students were asked: How often have you observed other students smoking cigarettes at your school? Responses ranged from 1 = never to 5 = every day. A higher score represents students perceiving a high frequency of other students using drugs at school. This is a unidimensional scale, with internal reliability, alpha = 0.83.Parental monitoring. This is a 7-item scale exploring how much or how frequently parents are aware or know about students' activities in the school, with friends, and after school. Responses ranged from 1 = little/never to 4/5 = very much/always. A higher score means that parents are well aware of where and with whom the students are spending time .Maltreatment. This is a 6-item scale exploring maltreatment experience at home. For example, the students were asked how often during the last year, they have experienced insults or threats from parents or any other adult at home. Answers ranged from 1 = never to 5 = every day. A higher score means higher frequencies of maltreatment experiences at home. This is a unidimensional scale, with internal reliability, alpha = 0.81.We asked students if they have perceived or experienced some of the following actions in their neighborhood, specifically (1) People selling drugs; (2) Fights in the streets; (3) People using drugs in the streets; (4) Assaults; (5) People carrying guns; (6) Bullet sounds; and (7) Unsafe places for walking. Responses for each item range from 1 = never to 5 = every day. This variable may be considered as related to the social disorganization theory , school co-educational status , number of teachers per school , school denomination , and co-educational status .Complete data analyses, using listwise deletion, were performed. General descriptive statistics were used to characterize the sample. The association between variables was assessed using multilevel logistic regression models for each substance. Multilevel analysis is recommended when data comes from hierarchical levels. In this study, the students belonged to schools where they share context; therefore, we expected the same degree of similarity between their behaviors. In other words, observations are not completely independent of one another  students from 754 schools was collected. The mean age was 14.9 (SD = 1.8) years old. Students were attending grades 7–12 . Conversely, students who were not exposed to any experience of bullying had the lowest substance use rates.Regarding the school domain, bully-victims showed the lowest school policy awareness, school bonding, school membership, and perceived the most of peer drug use at school.Additionally, bully-victims reported low level of parental monitoring, and high levels of maltreatment and community disorganization (see Table The full model had a good fit (C-statistic = 0.77). Regarding bullying experience, students who had any kind of bullying experience, compared with students with no bullying experience, had a higher risk for smoking, especially among those who bullied others. The risks were similar between victims  and bystanders , and between bullies  and bully-victims  , independent of whether they attended a mixed or only-girls school. In the school domain, higher academic achievement , stronger sense of school membership , and a good sense of school bonding  reduced the risk for smoking. On the contrary, students who perceived more using, passing, or selling drugs at school had a higher risk. In the family domain, parental monitoring seemed to reduce the risk for smoking . However, reporting a higher frequency of maltreatment actions at home increased the risks . The perception of actions of violence and community disorganization also increased the risk for smoking.At the school level, the only factor associated with smoking was the number of teachers per school. The higher the number, the lower the risk for smoking .The full model had a good fit (C-statistic = 0.78). Regarding bullying experience, students who had any kind of bullying experience, compared with students with no bullying experience, had a higher risk for drinking; and the risk was highest among those who bullied others. The risks were similar between victims  and bystanders  and between bullies  and bully-victims  . Those students who had good academic performance  and had a high sense of school membership  had a reduced risk for drinking. Parental monitoring also reduced the risk of drinking . On the contrary, students who reported higher acts of maltreatment at home  and/or a higher perception of violent or crime behavior near their homes , demonstrated an increased risk of drinking.At the school level, subsidized  and private  schools had a higher risk for drinking. Finally, students attending religious, non-Catholic schools had a reduced risk for drinking.The full model had a good fit (C-statistic = 0.81). Regarding the experience of bullying, those students who participated as bystanders , bullies , and bully-victims  had an increased risk for cannabis use. Victims had a higher risk, but not at a significant level , and those who had a strong sense of school membership  displayed a lower risk. Parental monitoring was also a strong variable associated with a reduced risk for cannabis use . Conversely, those students who reported maltreatment at home had a higher risk . Finally, the higher the perception of crime and drug use in the community, the higher the risk for cannabis use .At the school level, students attending subsidized  and private  schools appeared to have a higher risk for cannabis use than those students attending state schools.We present the results of one of the largest studies in Latin America exploring the association between the experience of bullying actions and substance use, among a representative sample of adolescents in Chile. We could control this association for several individual-level and school-level variables, using multilevel, multivariable logistic regression models.Firstly, we found that there is significant inter-school variability for each substance use explored, where cannabis use was the behavior with the highest variance explained by schools (11.4%). After controlling for several variables , the variance explained by school was reduced. However, there was still unexplained variance, which could be due to other unmeasured variables, such as teaching quality, quality of management by school authorities, the existence of preventive programs, and enforcement of norms. When we compared the contribution to the school effect between individual- and school-level variables, most of the reduction of the inter-school variability is due to pupil composition , rather than school variables.We have found that most students (85%) had played some part in bullying dynamics. As it has been found elsewhere, victimization was higher than being a bully  had a higher risk for substance use. There is evidence that bullies, especially those who are also victims (bully-victims), are in a higher risk for cigarette, alcohol, and cannabis use, which confirms results found in other studies  or having suffered domestic violence. Maltreatment at home had a robust influence on smoking, drinking or cannabis use in our study. The same is true for community disorganization which included the perception of violence in the neighborhood. Similar findings have been found elsewhere  may have a positive effect on another behavior, such as decreasing substance use. Of interest is the association found between a stronger sense of school membership and a reduced risk for substance use. This variable seems to have impacts on several other domains, such as for instance, academic performance.This study shows that bullies and bully-victims have a higher risk for cigarette, alcohol, and cannabis use. Individuals that were victims also demonstrated a high risk for cigarette and alcohol use; and bystanders displayed a high risk for cigarette, alcohol, and cannabis use. This is one of the few studies exploring the association between witnesses of bullying and substance use. These findings add new insights to the study of the co-occurrence of bullying and substance use. However, the mechanisms involved in these associations remain to be elucidated. We have provided good evidence that bystanders are an at-risk group, and this information should be taken into account when designing preventive interventions. Additionally, we have identified other factors related to substance use, such as school membership and parental monitoring, which can be considered in preventive interventions. Future research may also consider the potential benefits of implementing a preventive intervention to address one of these problems  and to study the effect on other problems .JG conceived the study, performed the statistical analyses, and drafted the manuscript. BT, DV, CR, CS, and RA participated in the interpretation of the data and contributed to the writing. EV participated in the design and coordination of the study, data collection and contributed to the writing. All authors read and approved the final manuscript.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


Many chronic fatigue syndrome (CFS) and fibromyalgia syndrome (FMS) patients (35–68%) use nutritional supplements, while it is unclear whether deficiencies in vitamins and minerals contribute to symptoms in these patients. Objectives were (1) to determine vitamin and mineral status in CFS and FMS patients as compared to healthy controls; (2) to investigate the association between vitamin and mineral status and clinical parameters, including symptom severity and quality of life; and (3) to determine the effect of supplementation on clinical parameters.st 1994 for CFS patients and 1990 for FMS patients till March 1st 2017 were included. Articles were included if the status of one or more vitamins or minerals were reported, or an intervention concerning vitamins or minerals was performed. Two reviewers independently extracted data and assessed the risk of bias.The databases PubMed, EMBASE, Web of Knowledge, and PsycINFO were searched for eligible studies. Articles published from January 1A total of 5 RCTs and 40 observational studies were included in the qualitative synthesis, of which 27 studies were included in the meta-analyses. Circulating concentrations of vitamin E were lower in patients compared to controls (pooled standardized mean difference (SMD): -1.57, 95%CI: -3.09, -0.05; p = .042). However, this difference was not present when restricting the analyses to the subgroup of studies with high quality scores. Poor study quality and a substantial heterogeneity in most studies was found. No vitamins or minerals have been repeatedly or consistently linked to clinical parameters. In addition, RCTs testing supplements containing these vitamins and/or minerals did not result in clinical improvements.Little evidence was found to support the hypothesis that vitamin and mineral deficiencies play a role in the pathophysiology of CFS and FMS, and that the use of supplements is effective in these patients.http://www.crd.york.ac.uk/PROSPERO/display_record.asp?ID=CRD42015032528.Study methods were documented in an international prospective register of systematic reviews (PROSPERO) protocol, registration number:  Chronic fatigue syndrome (CFS) and fibromyalgia syndrome (FMS) are syndromes of unknown origin. The core symptom of CFS is profound disabling fatigue , whereasVitamin and mineral deficiencies may play a role in the pathophysiology of both CFS and FMS, although mechanisms behind this hypothesis are not entirely clear ,8. In adRecently, a review investigating hypovitaminosis D in both chronic pain and FMS patients showed that these patients were at significantly higher risk of hypovitaminosis D than healthy controls . Unfortuhttp://www.crd.york.ac.uk/PROSPERO/display_record.asp?ID=CRD42015032528.We followed the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) guidelines  19]. Pr. Pr19]. st 1994 and 1990, for CFS and FMS respectively, and March 1st 2017 were included. We focused on the most recent diagnostic guidelines, namely the International Center of Disease Control (CDC) diagnostic criteria for CFS that was established in 1994 [The databases PubMed, EMBASE, Web of Knowledge, and PsycINFO were systematically searched. Articles published between January 1Title and abstract were screened by two independent reviewers (M.L.J. and I.M.) for the following criteria: (1) CFS or FMS patients; (2) vitamin or mineral status; and (3) study design. Studies which were in agreement with the eligibility criteria were retrieved as full text. Discrepancies between the two researchers were resolved by consensus, and when needed a third assessor was consulted (J.G.M.R.). Reasons for exclusion and percentage of agreement, as Cohen’s kappa, between the assessors were documented.Participants of the included studies had to be adults i.e. ≥18 years) suffering from CFS or FMS according to the official diagnostic criteria 8 years s. StudiesTwo reviewers (M.L.J. and I.M.) independently extracted data and assessed the risk of bias for each study. The first ten articles were screened together to pilot the data extraction and risk of bias form. Reasons for exclusion and percentage of agreement between the assessors were documented.From the included articles, the following information was extracted: name first author, publication year, type FSS, number and age of the participants, and vitamin or mineral status. In addition, data on smoking habits or alcohol use, diet (and assessment tool used), BMI , physical activity (assessment tool), socioeconomic status, ethnicity, severity of illness (assessment tool), duration of illness, co-morbidities (somatic and psychiatric), medication use, clinical parameters including symptom severity and quality of life, and in case of RCTs the relevant co-intervention(s) were also extracted.To assess quality of RCTs, the Cochrane Collaboration’s tool for assessing risk of bias was employed . For obs2) value. Publication bias was inspected visually by a funnel plot, and an Egger’s test was conducted to quantify funnel plot asymmetry [We first constructed an overview of available data on the different vitamins and minerals. Characteristics of the included studies were systematically listed to generate a clear overview of the current literature on vitamins and minerals in CFS and FMS patients. For those vitamins and minerals with more than five studies available, we did quantitative syntheses on aggregated data. For these syntheses, data was pooled with the random effects model of meta-analysis, using Stata statistical software, version 14 . To allow pooling across studies that used different outcomes of vitamin or mineral plasma or serum levels, we calculated the standardized mean difference (SMD). For proportions of deficiencies, the odds ratio (OR) was calculated and pooled. Subsequently, the SMD and OR for each study were weighted by their inverse variance and the corresponding 95%CI were calculated. The existence of heterogeneity among studies was assessed by Q-tests, and the degree of the heterogeneity was quantified by calculating the I-squared  , vitaminAll studies that investigated vitamin B12 (n = 1) , folic aVitamin C, vitamin D, vitamin D deficiency (<20ng/ml), vitamin E , and theSubgroup analyses were performed including studies with more than half the maximum study quality score , if more than three studies with a sufficient quality score were available. The additional analysis was not possible for magnesium, since only two studies achieved more than half of the maximum quality score. No differences in circulating concentrations of vitamin C  ,42,43,672 = 13.3% versus 88.5% in the overall analysis including CFS patients), indicating a high consistency of studies’ results. The significant difference in circulating concentrations of vitamin E between patients and controls disappeared when the single CFS study was excluded .Subgroup analyses were performed separately for the syndromes, when more than three studies were available per syndrome. Since vitamin D, vitamin D deficiency and calcium were only determined in FMS patients, additional subgroup analyses were possible for vitamin C, vitamin E and magnesium. No statistically significant difference between patients and controls was found in the three studies investigating circulating concentrations of vitamin C in FMS patients . However, the heterogeneity was substantially lower  and confounding factors. Potential confounders were assessed in about half of the studies, but almost no studies adjusted their analyses for potential confounders. Consequently, the results of the current review may be affected by the methodological weaknesses that are accompanied by the observational study designs. Second, quality assessment revealed a poor study quality in the majority of studies. This demonstrates that substantial improvements can be made in terms of study quality, especially in specification of in- and exclusion criteria, presenting disease characteristics of the participants, making use of validated methods to assess vitamin and mineral status, to perform the vitamin and mineral assessments in duplicate, and, as mentioned earlier, to adjust analyses for potential confounders. Furthermore, a quality issue in research on CFS and FMS patients is that of careful selection of control groups. Our quality assessment showed that many included studies fell short because of the selection of the controls, which could result in inaccurate study results. Third, a problem that affects the validity of meta-analyses is the presence of publication bias. Funnel plots indicated the absence of publication bias in the majority of the meta-analyses. Trimming was performed among the calcium studies, and two “missing” studies were added, while no significant funnel plot asymmetry was present. However, trimming was performed in the statistically significant areas, which argues against the presence of publication bias. Although Egger’s test is preferred for more than 10 studies, it revealed significant funnel plot asymmetry in vitamin E, while no trimming was performed. It is therefore possible that the significant outcomes of vitamin E in patients are influenced by publication bias. Lastly, a substantial to considerable heterogeneity in most studies was found, which makes it difficult to reach a final conclusion about vitamin status in CFS and FMS patients.This review reveals that very few RCTs have investigated the effect of vitamin and mineral supplementation versus placebo in CFS and FMS patients. Most published RCTs found no treatment effect of vitamin and mineral supplementation on clinical parameters. So, the evidence for beneficial effects of supplementation in CFS and FMS patients is not proportional to the large quantity of supplements that are used by these patients. Nevertheless, the industry of vitamin and minerals supplements is increasing, for example, Americans spend an estimated $36.7 billion each year on supplements . This isWe conclude that there is little evidence to support the hypothesis that vitamin and mineral deficiencies play a role in the pathophysiology of both CFS and FMS. Furthermore, the current literature on vitamins and minerals in CFS and FMS is of poor quality and stresses the need for well-performed intervention research, and large population-based and age-matched prospective studies in CFS and FMS, in order to gain more insight in the role of vitamins and minerals in the pathophysiology of CFS and FMS. According to our results, potential vitamins and minerals that should be further examined include vitamin A and vitamin E.S1 Table(DOCX)Click here for additional data file.S1 Appendix(DOCX)Click here for additional data file.S2 Appendix(DOCX)Click here for additional data file.


During the past two decades, several scientific publications from different countries have shown how oral health in the population varies with social determinants. The aim of the present study was to explore the relationship between self-reported oral and general health in relation to different measures of socioeconomic position.n = 3500, mean age 53.4 years, 53.1% women). The response rate was 49.7%. Subjects were interviewed by telephone, using a questionnaire including items on self-reported oral and general health, socioeconomic position and lifestyle.Data were collected from a randomly selected sample of the adult population in Sweden  and with oral health (OR 1.76) if having an income below SEK 200,000 per year. Similar results were found in multivariate analyses controlling for age, gender and lifestyle variables.For adults, there are clear socioeconomic gradients in self-reported oral and general health, irrespective of different socioeconomic measures. Action is needed to ensure greater equity of oral and general health. During the past two decades, several scientific publications from different countries have shown how oral health in the population varies with social determinants –13. The n = 3500, giving a participation rate of 49.7%. Due to the non-response of the non-participants, a comparison between participants vs. non-participants was not possible to conduct. However, when comparing with official data for the general Swedish population  certain differences were found . Participants in the survey were somewhat older (53 vs. 49 years of age), proportionally more women were included in the study (53.1% vs. 50.5%) and fewer foreign-born (10.1% vs. 18.0%). The Regional Ethical Review Board in Gothenburg, Sweden approved of the study (Reg. No. 801–12).The present survey included a random sample of the adult population in Sweden. A telemarketing company, TNS SIFO, which is a Swedish company that performs public opinion and market surveys, was responsible for the sample selection and the interviews with the individuals by telephone. The participants were randomly selected from the SPAR register in Sweden. The SPAR  includes all individuals who are registered as being resident in Sweden. The exclusion criterion was individuals who did not speak and/or understand Swedish. The number of adults (aged ≥19 years) included in the sample was (i)Self-perceived health: How good is your general health? (five response options: poor (1), fair, good, very good, excellent (5)). The scale was dichotomised into poor/fair (Poor) vs. good/very good/excellent (Good); How good is your oral health? , fair, good, very good (4). The scale was dichotomised into poor/fair (Poor) vs. good/very good (Good).(ii)SEP variables, such as gender, marital status (married/cohabiting vs. single), income in SEK  for the whole household (SEK 10 ~ 1 euro), level of education , however in the multivariate analyses the categories university and Masters or PhD degree were collapsed into one category (university), estimated financial resources for unforeseen expenditure , however in the multivariate analyses the categories “no mostly not” and “no never” were merged into one category. Ethnicity was categorised as foreign-born or Swedish-born, including the Nordic countries, and housing as rented flat, owner-occupied flat, or own house.(iii)Lifestyle variables: How much do you exercise  during your leisure time? , a little, once a week, twice a week or more, intensive exercise at least twice a week (5)); smoking .A questionnaire with items concerning demography, self-rated oral and general health, education, income, dental care behaviour, and lifestyle issues was used in the study. There were a total of 39 questions; however, the questions pertaining to self-rated health and socioeconomic position (SEP) were the focus of the analysis. The following variables were used in the analysis:N = 502 answers are missing. When multiple comparisons were performed, Bonferroni corrections were applied.The descriptive analysis included frequencies, measures of central tendency (means), and variability (standard deviation). Bivariate analyses were performed using the t test, the Mann–Whitney test, the chi-square test, and logistic regression. Multivariate logistic regression analysis (MLR) was used with the self-rated health variables as the dependent variables and the socioeconomic indicators as the independent variables. Two different models were applied to the MLR; Model 1 included age and gender as covariates, and in Model 2, the analysis also included the lifestyle factors of smoking and physical activity. Model evaluations were performed using the Hosmer-Lemeshow and Nagelkerke test statistics. The pre-selected level of significance was α = 0.05. Due to some missing responses, the numbers of observations differ in the analysis. Thus, missing answers in the presented variables range between 0 and 52, except for the variable of income, where Tables The proportions of poor health relative to the SEP variables are shown in Tables . Another pattern in the data is that the odds ratios are of a higher magnitude in all categories for poor general health than for poor oral health. The SEP variable showing the steepest gradient is the income measure with an odds ratio of 3.95, indicating an almost four times higher risk of reporting poor general health if having an income below SEK 200,000 per year. For poor oral health, the highest odds ratio (2.61) was found when individuals reported no or little possibility of obtaining a sum of SEK 15,000 in one week.Obvious patterns were revealed Table  when theThe multivariable models are shown in Tables This cross-sectional epidemiological survey presents data from Sweden analysing how self-reported general and oral health are associated with different types of socioeconomic measures. The sample consisted of randomly chosen adult individuals who were asked to answer a battery of questions during a telephone interview. The main findings were that all SEP variables—education, income, the possibility of obtaining a given sum of money in one week, and the participants’ type of housing—had (i) a similar strength of association with the respective outcome, and (ii) showed a gradient relationship with the outcome. The lower the educational level, the lower the income, the smaller the chance of obtaining a large sum of money, and the poorer the housing situation, the higher the risk of reporting poor general and oral health. These findings were found in bivariate as well as in multivariate analyses taking into account age, gender and lifestyle factors important for health. However, income and the ability of obtaining a given sum of money in one week, were the SEP measures that showed the strongest associations with the two health outcomes.The main objective of the study was to investigate gradients in self-reported health relative to different socioeconomic measures and to evaluate the strength of the associations depending upon the influence of other important risk factors for poor health. Over the past three to four decades, the scientific literature has shown typical strong correlations between health—both self-reported and objective health measures—and a plethora of indices capturing socioeconomic status , 5, 8. TThe findings from this study indicate that the four different SEP variables have similar patterns with regard to gradients in the ORs. However, the variables most clearly associated with material resources in a short-term perspective; i.e., income and ability to obtain a large sum of money in one week, showed the strongest risk ratios with respect both to general and oral health. This result may be indicative of a previously revealed association between financial resources and health status . It coulIn the statistical modelling, we reduced the number of categories in some of the SEP variables, such as education and the ability to obtain money in one week, due to few responses in some categories. However, when all the categories were used in descriptive tables, the differences in self-reported poor general and oral health still revealed the same gradient pattern. For example, in the education variable, individuals with masters or PhD degrees were in the highest category and there were significantly fewer participants reporting poor health in this category, also compared with individuals with a university degree but without a research degree. Similar results have been reported by Robert Erikson , where tThis survey comprising a random sample of adult Swedish individuals has some weaknesses and strengths. The proportion of non-respondents was high (50.3%), and the mean age, proportion of women and Swedish-born in the sample was significantly higher compared with the official Swedish data. It may be argued that the steepness of the gradients and the strength of the odds ratios would be underestimated, as has been revealed in other large-scale epidemiological surveys in Sweden . Thus, aIn conclusion, the results of this study show that: (i) there are socioeconomic gradients in self-reported health, both general and oral, irrespective of different socioeconomic measures; (ii) the slope of the odds ratios is similar for general and oral health and the presented socioeconomic measures, and (iii) there are significant and important differences in perceived health among adult individuals living in Sweden. The obvious interpretation of these results is that action is needed for better health and health care through upstream and downstream activities, both prevention and promotion, to ensure greater equity of health.


The standard APD battery identified a larger proportion of participants as having APD, than an attention battery identified as having Attention Deficits (ADs). The proposed APD battery excluding AD cases did not have a significantly different diagnosis proportion than the standard APD battery. Finally, the newly proposed diagnostic battery, identifying an inattentive subtype of APD, identified five children who would have otherwise been considered not having ADs. The findings show that a subgroup of children with APD demonstrates underlying sustained and divided attention deficits. Attention deficits in children with APD appear to be centred around the auditory modality but further examination of types of attention in both modalities is required. Revising diagnostic criteria to incorporate attention tests and the inattentive type of APD in the test battery, provides additional useful data to clinicians to ensure careful interpretation of APD assessments.Measures of attention have been found to correlate with specific auditory processing tests in samples of children suspected of Auditory Processing Disorder (APD), but these relationships have not been adequately investigated. Despite evidence linking auditory attention and deficits/symptoms of APD, measures of attention are not routinely used in APD diagnostic protocols. The aim of the study was to examine the relationship between auditory and visual attention tests and auditory processing tests in children with APD and to assess whether a proposed diagnostic protocol for APD, including measures of attention, could provide useful information for APD management. A pilot study including 27 children, aged 7–11 years, referred for APD assessment was conducted. The validated test of everyday attention for children, with visual and auditory attention tasks, the listening in spatialized noise sentences test, the children's communication checklist questionnaire and tests from a standard APD diagnostic test battery were administered. Pearson's partial correlation analysis examining the relationship between these tests and Cochrane's Q test analysis comparing proportions of diagnosis under each proposed battery were conducted. Divided auditory and divided auditory-visual attention strongly correlated with the dichotic digits test,  Auditory Processing Disorder (APD) is characterised by normal peripheral hearing, but abnormal processing of auditory information within the central auditory nervous system  and AP tests  have not been sufficiently investigated. It is hypothesised that in addition to Sus-AA correlating with specific AP measures, as previously shown  test was set. This excluded three children who were below that threshold. Thus, the study sample comprised 27 UK children (9 girls and 18 boys) suspected of APD, aged 7–11 years (mean age 9 years 7 months). All children had English as first language and three were bilingual. None of the children previously used a Frequency Modulation (FM) system or auditory training. None met diagnostic criteria for ADHD. All children and parents signed assent and consent forms, respectively. The Bloomsbury Research Ethics Committee approved the study.Initially, 85 invitation letters were sent to families of children with parental reports of Speech-in-Noise (SiN) difficulties and with referrals by audiologists for APD assessment at Great Ormond Street Hospital in London on suspicion of AP deficits. Thirty replies were received and an inclusion criterion of at least a All children were tested for APD following the standard APD diagnostic procedure. First, pure tone audiogram and tympanogram/ acoustic reflexes were used to confirm normal peripheral hearing. Children completed AP tests at Great Ormond Street Hospital as per the clinic's APD protocol. The AP tests comprised the Dichotic Digits Test (DDT), Gaps-in-Noise (GiN), Frequency Pattern Test (FPT), and SCAN-3 Auditory Figure Ground (AFG) +8 dB and 0 dB. The DDT presents 20 stimuli  and the child needs to repeat back the numbers from both ears  Scale of Ability, the Test of Everyday Attention for Children (TEACh) and the Listening in Spatialized Noise-Sentences (LiSN-S) test  connected to two ALTEC LANSING speakers was used with the sound level measured at ear level from the child's seat at 60 dB sound pressure level with a calibrated Casella CEL-450 sound level meter. The LiSN-S test ran on a ProBook HP laptop via the Phonak soundcard and Sennheiser HD125 headphones, which kept competing speech (either coming from the side or front) at a constant level of 55 dB sound pressure level and frontal target sentences at an initial 62 dB sound pressure level, then adjusted 2 dB up or down if at least 50% of the sentence words were incorrect or correct, respectively   identified children as having APD when at least two of the AP test scores were 2 Standard Deviations (SDs) below the mean or when only one test was 3 SDs below the mean  and typical ear-canal volumes (between 0.4 and 1.0 cm3)   were met. Gender and performance in the WNV test were controlled in the model, while age was not a covariate as all scores were converted into z scores. Results are outlined in Table Twenty of the 27 children met criteria for APD diagnosis under the standard APD battery described previously. Pearson's partial correlation analysis was conducted on these 20 children with confirmed APD to examine the correlations between attention and AP tests in this subgroup. Scores from the right and left ear of the GiN and DDT were averaged as they were not significantly different. All assumptions of linearity, absence of outliers and normality of distribution  = 0.76, p < 0.001 and r(18) = 0.51, p = 0.03, respectively, while Div-AA correlated with Div-AVA, r(18) = 0.53, p = 0.02. Divided auditory attention and Div-AVA both demonstrated strong associations with DDT scores, r(13) = 0.68, p = 0.01, and r(13) = 0.76, p = 0.003, respectively. These two correlations are also graphically detailed in Figure There were some intercorrelations between some TEACh conditions. The TEACh Sus-AA condition correlated with the Div-AA and Div-AVA conditions, r8 = 0.76,r(17) = 0.53, p = 0.03, and r(17) = 0.66, p = 0.004, respectively. There were also some associations between same-test conditions. Finally, LiSN-S High-cue associated with LiSN-S Total Advantage, r(18) = 0.93, p < 0.001, whereas the CCC-2 SLI score correlated with the CCC-2 PLI score, r(17) = 0.59, p = 0.01.Sustained auditory attention and Div-AA both correlated with the CCC-2 SLI score, n ≥ 4 and sample size, n, multiplied by number of related groups, k, nk ≥ 24). The diagnostic yield between these three Batteries was significantly different, χ2(2) = 11.200, p = 0.004. This was followed up by pairwise comparisons using Dunn's procedure with a Bonferroni correction for multiple comparisons with adjusted p-values reported. The APD Battery, with 74% diagnoses, was significantly higher than the AD Battery with 30% diagnoses, p = 0.003. The other comparison that was examined between the diagnostic yield of the APD Battery and the APD (without ADs) Battery (44%) was non-significant.Table Finally, as seen in Table r = 0.53, p = 0.06. Given the small sample size of the study and the wide confidence interval of the correlation coefficient, it is not possible to draw meaningful conclusions about the relationship between Sus-AA and DDT from the data. None of the other AP tests  showed significant correlations with Sus-AA, which agrees with the literature on APD-suspected children  and DDT in children diagnosed with APD. Divided auditory attention tasks and DDT tasks both require listening to two separate streams simultaneously and it is thus not surprising to observe such associations between their test scores. This means that during dichotic listening, divided attention might be used and/or similarly, in Div-AA tasks binaural integration skills could be employed. It could also be argued that the use of the DDT for diagnosing APD should be reconsidered, granted the relationship it has with divided attention tasks. It is not possible, though, through this analysis to determine to what extent the deficits in attention (divided and sustained) cause poorer performance in the AP measures or vice versa and how the separate deficits contribute to the problems the child has in real life. Bigger sample sized studies that would include tests that control for task attention demands and complexity of auditory test stimuli would be better suited to provide answers to these questions. These findings serve as a basis for future research on divided attention and APD.It can be observed in Table Finally, the lack of association between divided attention tasks and the rest of the AP tests shows that these AP tests are not strongly influenced by attention  and that they are good tools to use in an APD diagnostic battery. The DDT, revealing strong correlations with divided attention, should be reconsidered as an APD diagnostic tool, given the possible influence it might have from attention. Nevertheless, this argument needs further investigation, primarily to determine the direction of causality between these two measures. Alternatively, there is some evidence that divided attention could be linked to the function of the splenium of the corpus callosum and that children with ADHD have smaller splenium  might not be the problem with APD children. However, further investigation of this relationship and view is required, since the sample size was small and the confidence interval of the correlation coefficient wide. Besides, 10% of the APD diagnosed children had rates of disordered performance in the visual attention test. While this is comparatively lower than the rates in the three tests of auditory attention (which ranged from 30 to 55%), it still exceeds the 2% disordered threshold in the test's normative sample  also received APD diagnosis  were identified. While all five children would have been diagnosed as having APD in the absence of ADs (using the APD without ADs Battery), and thus not receive any recommendations for addressing attention problems, they do present with difficulties in one attention test that would have been otherwise gone undetected. Thus, the usefulness of the i-APD Battery is to identify cases of children with APD that appear to have an inattentive type of the disorder, but not broader auditory attention difficulties. Further examining the tests in which these children underperformed, it was observed that 4 out of 5 of them had deficits in a divided attention task and one of them in Sus-AA. Hence, a small subgroup of children with APD  might have a specific underlying difficulty in performing divided attention tasks. Therefore, this group of children, in addition to the standard APD management strategies mentioned above, could receive cognitive training targeting divided attention , as this was not possible in the present study due to the reduced group sizes. Finally, as the study used the TEACh test, which lacks sub-tests that measure both modalities in the different types of attention, further examination of sustained, selective and divided attention in the auditory and visual modality should take place. This could help confirm the domain general attention problems children with APD face.The findings of the study have implications for clinical practice and APD diagnosis, as the proposed diagnostic procedure takes into account specific attentional deficits in order to classify children in better-defined categories. Incorporating attention tests in the APD diagnostic test battery adds information of value in terms of the potential clinical management of these children by identifying cases with predominantly inattentive type of APD, as well as children with general ADs. Results from the correlation analysis demonstrate significant correlations between divided attention measures and the DDT, suggesting reconsideration of the use of DDT for APD diagnosis. However, an undetermined direction of causality in this relationship stresses the need for further research to clarify this relationship. The non-correlation of the visual attention task with auditory processing tests and the fact that only two cases had Sel-VA deficits, suggests that the majority of APD children are not at risk of broader attentional problems, but instead face major deficits in auditory attentions tasks (sustained and auditory). Larger sample-sized trials including both visual and auditory attention tests of the same type are needed to cross-validate these findings and to clarify the relationship between attention and APD, which could in turn better inform choice of appropriate management strategies to address listening difficulties in real-life environments.The study received ethics approval from the London–Bloomsbury Research Ethics Committee (REC). REC reference number: 14/LO/1509. Written and verbal explanation of the research aims was given to both parents and their participating children. They both had the opportunity to ask questions about the project. Assent forms were signed by children in the presence of their parents and parents in turn signed consent forms.GS: Conception, design and draft of the work, analysis, acquisition and interpretation of data, revising the work critically. V-MI: Contribution to the conception and design of the work, interpretation of data, revising the work critically. LE: Contribution to the design and draft of the work, revising the work critically. TS: Acquisition of data, revising the work critically. D-EB: Conception, design and draft of the work, acquisition and interpretation of data, revising the work critically. GS, V-MI, LE, TS, and D-EB: Final approval of the version to be published, agreement to be accountable for all aspects of the work in ensuring that questions related to the accuracy or integrity of any part of the work are appropriately investigated and resolved.The research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest. The authors declare that this study received funding from GN ReSound and Action Medical Research. Neither the funder nor the supporter were involved in the study design or collection, analysis, interpretation of the data or writing of the report. The reviewer, AL, and handling Editor declared their shared affiliation.


The prognostic nutritional index (PNI) has been reported to correlate with the prognosis in patients with various malignancies. We performed a meta-analysis to determine the predictive potential of PNI in digestive system cancers. Twenty-three studies with a total of 7,384 patients suffering from digestive system carcinomas were involved in this meta-analysis. A lower PNI was significantly associated with the shorter overall survival (OS) , the poorer disease-free survival (DFS) , and the higher rate of post-operative complications . In conclusion, PNI was allowed to function as an efficient indicator for the prognosis of patients with digestive system carcinomas. According to the International Classification of Diseases for Oncology, digestive system carcinomas include esophageal, gastric, hepatocellular, colorectal and pancreatic carcinomas [childhood and adolescent carcinomas are classified according to the International Classification of Childhood Cancer (ICCC)] . Despiteet al. in 1980 [et al. in 1984 [Several nutritional and immunological indicators, such as PNI, platelet-to-lymphocyte ratio (PLR), neutrophil-to-lymphocyte ratio (NLR), and the Glasgow prognostic score (GPS), have effeciently functioned as the assessable factors for uncovering the prognosis of cancers patients and for procuring the optimal therapy –12. The  in 1980 , which w in 1984 . Recent  in 1984 , 16, 17. in 1984 –15. PNI As shown in Figure P values or the relative data from other 6 studies.A total of 7,384 participants were enrolled with the sample sizes of the individual studies ranging from 80 to 1321 (mean size: 321). 22 of 23 studies originated from Asian countries , and one from Scotland. Five patterns of carcinomas were analysed in this systematic review: gastric cancer (7 studies), colorectal cancer (7 studies), hepatocellular carcinoma (4 studies), pancreatic cancer (3 studies), and esophageal carcinoma (2 studies). Seventeen studies investigated the prognostic significance of PNI with respect to post-operative outcomes, while 6 studies did not disclose whether patients had received surgery or what percentage of patients received surgery. The cut-off value of the PNI ranged from 40 to 49.7 with 43.5% of the studies (10 of 23) having a cut-off value set at 45. In accordance with the respective cut-off values, all studies divided patients into two groups: patients with low-PNI and high-PNI group. OS, DFS/PFS, cancer-specific survival (CSS), and post-operative complications were the main variances in the enrolled studies. There were 30 HRs/ORs extracted from 23 studies, including 22 for OS, 3 for DFS, 2 for CSS, and 3 for post-operative complications. Most of HRs/ORs and 95% CIs of the estimates could be directly obtained from 17 studies, while the small part of the estimates were indirectly calculated using the HRs and I2 = 50.3%; P = 0.004). Then, we stratified the studies into subgroups according to human cancer type, sample size, region, cut-off value, and paper quality score to explore the differences between the subgroups in order to elaborate the prognostic potential of PNI for OS   Figure .I2 = 61.3%, P = 0.024) and Japanese subgroups had significant heterogeneity . The one study originated from Scotland demonstrated a similar HR in colorectal carcinoma  [For the studies originated from the three Asian countries, the pooled HRs were the following: . Chinese (99–1.79)   displayed significant heterogeneity while the former subgroup didn't   in the subgroup of studies with cut-off values lower than 45, while the pooled HR was 1.74 (95% CI 1.54–1.97) in the subgroup of studies with cut-off values higher than 45. Latter subgroup  Figure .P value (P = 0.001) and the P value of Egger's test was 0.000 . Because of the limited number of included studies, we didn't perform subgroup analysis, meta-regression, sensitivity analysis, Begg's test and Egger's test.There were three studies with a total of 1009 patients, three studies with a total of 979 patients, and two studies with a total of 749 patients analysed the associations between PNI and post-operative complications –7, DFS , 28, 30 This meta-analysis study aimed to investigate the association between PNI and the prognosis of patients with digestive system carcinomas. 23 studies involving five types of digestive system cancers were enrolled in our meta-analysis. We found that a lower PNI played a negative role as an independent prognostic factor for patients’ survival: shorter OS and DFS, as well as the appearance of post-operative complications of digestive system cancers, especially in gastric carcinomas, colorectal tumours, and hepatocellular cancers. The followings are some possible explanations about the association between low PNI and poor prognosis in cancer patients. Firstly, PNI is decided by the serum albumin concentration and lymphocyte count in the peripheral, both of which are obviously correlated with the prognosis of cancer patients. PNI reflexes the healthy condition of cancer patients. Secondly, poor nutritional status may make patients delay the surgery or adjuvant therapy, even be untreatable for disease. Therefore, this may partly explain the association between low PNI and poor prognosis of patients with cancers. Because of these direct or indirect clues, a low PNI is related to the prognosis of cancer patients.P values of two tests, we considered that publication bias existed in this meta-analysis. Although we detected the heterogeneity in the meta-analysis, the sensitivity analysis ensured the stability of the results. There might be several factors being responsible for underlying heterogeneity, such as the patients from different regions, the wide range of publication date, the uneven quality of the papers, the development of detection techniques, the improvement of surgical efficacy and safety, the statistical approach for extrapolating HRs, and the difference in cancer staging criteria. The cut-off value could be considered consistent in our study, but it was still required to verify that 45 was an appropriate cut-off value for our future study. Under these circumstances, individual study may have a considerable impact on publication bias and heterogeneity. Accordingly, more studies with larger sample sizes are required to further confirm this association.The Egger’s test and Begg’s test were carried out to assess publication bias. According to the This meta-analysis is the first systematic review concerning about the predictive value of PNI for patients with digestive system carcinomas, and circumvents the limitations of small sample sizes in the individual studies. 23 articles in our analysis included a total of 7,384 participants. Moreover, the data could almost be extracted directly from original articles reducing the approximation bias. Subgroup analysis, Egger’s test, Begg’s test and sensitivity analysis were applied to address this issue. However, there were certain limitations in our study. Firstly, most of the data originated from retrospective observational studies without the use of controls. Secondly, the studies eligible for this meta-analysis were inconsistent with respect to the staging outline of human cancers. Thirdly, all of the included studies almost were Asian origin, so it was unclear whether our findings extended and generalized to other regions of world as well. Thereby, the limitations urged to yield great promise for the addition of more studies with larger sample sizes and more nutritional factors. Moreover, we also sought to link the other nutritional factors with the prognosis of cancer patients, like PLR, NLR and GPS. It was reported that those factors all have negative effects on various cancers patients’ survival –42. ThouIn conclusion, we demonstrated that PNI was an independent predictive factor for OS, DFS, and post-operative complications. A lower PNI was indeed associated with a poorer prognosis in digestive system carcinomas. It does a favor to determine the optimal timing of surgery or choose the other appropriate individual therapy to adjust patients’ nutritional condition. It also can assist doctors to improve patients’ nutritional conditions by means of early and effective interventions to promote the survival outcomes.We performed a literature search in PubMed and the ISI Web of Science for articles with the relevance to the association between PNI and cancer patients. The following MeSH and free-text terms were used: ‘prognostic nutritional index’, ‘PNI’, ‘cancer’, ‘carcinoma’, ‘tumour’, and ‘neoplasm’. Subsequently, we checked the references of these papers in order to locate additional articles that could be included in the meta-analysis. We did not exclude articles based on language or publication date. The most recent article had been published on January 12, 2016.After the initial literature search, we handpicked the eligible studies according to the following criteria: (1) the investigations of the prognostic value of PNI in digestive system carcinomas; (2) availability of the necessary data, such as the OR or HR with 95% CIs, or sufficient data to calculate them; (3) availability to the full text; (4) written in English. Abstracts, meetings, and case reports were excluded.The quality of each paper was assessed using the Newcastle-Ottawa scale. Three aspects were evaluated: selection, comparability, and exposure. A paper can be awarded a maximum of one star for each numbered item within the Selection and Exposure categories. A maximum of two stars can be given for comparability. So, a paper can be awarded 9 stars at most, with more stars indicating higher quality.We divided the data that was extracted from each individual study into two categories. The first category contained the characteristics of the studies, including names of authors, publication year, region, tumour type, sample size, age of the patients, cut-off value, treatment, outcome measurements, and quality score. The second category contained the estimated ORs or HRs regarding the prognostic significance of PNI with respect to OS, DFS, CSS, and post-operative complications.P value or the relative data.Two methods described in previous studies , 44 wereI2), with P < 0.05 indicating significant difference and larger values of I2 indicating higher heterogeneity. Subgroup and sensitivity analyses were performed to reduce and explain the statistical heterogeneity when that was necessary in order to elaborate on the prognostic role of PNI in cancer patients. Graphical funnel plots were generated for visual inspection in order to qualitatively assess publication bias, while the Begg’s test and the Egger’s test were used for quantitatively determining the extent of publication bias [HRs of individual studies were obtained as described above and summarized as pooled HRs. The results presented as HRs and 95% CIs which were first calculated using the random-effects model in order to assess heterogeneity. If heterogeneity was not significant, the fixed-effects model (Mantel–Haenszel) was subsequently introduced . Heterogion bias , 47. All


Findings reveal that coercive power increases an antagonistic climate and enforced compliance, whereas legitimate power increases reason-based trust, a service climate, and voluntary cooperation. Unexpectedly, legitimate power is additionally having a negative effect on an antagonistic climate and a positive effect on enforced compliance; these findings lead to a modification of theoretical assumptions. However, solely reason-based trust, but not climate perceptions and motives, mediates the relationship between power and intended cooperation. Implications for theory and practice are discussed.The execution of coercive and legitimate power by an authority assures cooperation and prohibits free-riding. While coercive power can be comprised of severe punishment and strict monitoring, legitimate power covers expert, and informative procedures. The perception of these powers wielded by authorities stimulates specific cognitions: trust, relational climates, and motives. With four experiments, the single and combined impact of coercive and legitimate power on these processes and on intended cooperation of n In a community, contributions to public goods are often obligatory , but some individuals exploit the vulnerable system, refraining from participation, and consequently free ride , the relational climate (antagonistic and service climate), and motives for contribution . The study investigates how coercive power and legitimate power solely or in combination over perceptions of power influence trust in authorities, the climate between authorities and individuals, and the motives of cooperation. Additionally, it analyzes whether the cognitions such as trust, perceived relational climates or motives, mediate the relationship between power and intention to cooperate.In the remainder of this article, the impact of coercive power and legitimate power on cooperation, trust, relational climates, and motives are defined. Three laboratory experiments in the tax context and one online experiment in the insurance context are described, each assessing the impact of power. Finally, we discuss the results and identify their theoretical and practical implications for legislation and law enforcement.coercive power based on deterrence and legitimate power based on persuasion  to make cooperation easier and non-cooperation more difficult  majoring in several different disciplines participated on a voluntary basis and were paid according to their behavior in the experiment. As student populations are specifically naïve regarding experiences with tax authorities and our hypotheses, they specifically suit hypotheses testing in this context  or 7.66 USD (SD = 1.56), respectively].In the two conditions, the tax authority held either low or high levels of coercive power. A tax authority with low/high levels of coercive power was, for example, described as “… well-known for its lenient/hard sanctions.” After participants were introduced to the experimental set-up, a taxpayer's life was simulated using the software z-Tree  to 7 . Cronbach's α for the eight scales were excellent and can be found in Table In all treatment conditions, participants had to fill out a questionnaire. The questionnaire assessed implicit trust , reason-based trust , the antagonistic climate , the service climate , enforced compliance , and voluntary cooperation . As a manipulation check, we asked participants' perceptions of the tax authority's coercive power  and legitimate power  by adapting published scales from the organizational context  were also assessed.low) and high (cphigh) levels of perceptions regarding coercive power were in line with the manipulation  as factor variables and implicit and reason-based trust, antagonistic and service climate, and enforced compliance and voluntary cooperation as dependent variables  compared to when coercive power was low  rather than low  compared to low levels of coercive power  do not indicate a significant mediation.The mediator analysis revealed one indirect effect from coercive power to tax honesty intention, the relation was only found to be mediated by implicit trust  majoring in different fields participated on a voluntary basis and were paid based on their behavior in the experiment. Again, this population was selected because of their naivety regarding experiences with tax authorities  or 8.18 USD (SD = 1.76), respectively.The experimental design and procedure was similar to Study low: M = 3.16, SD = 1.05; lphigh: M = 4.89, SD = 1.16; Table high: M = 4.51, SD = 1.57) rather than low .The manipulation was successful as low and high levels of legitimate power induced perceptions according to the manipulation  as factor and implicit and reason-based trust, antagonistic and service climate and enforced compliance and voluntary cooperation as dependent variables  compared to when legitimate power was high . In line with the hypothesis, the perception of a service climate increased with legitimate power  compared to participants in the condition of low levels of legitimate power . In line with predictions, participants experiencing high levels of legitimate power reported higher levels of voluntary cooperation  than did participants experiencing low levels of legitimate power  compared to low levels of legitimate power  majoring in different disciplines participated in the experiment and were paid based on their behavior in the experiment.Analogous to Study and low or high levels of legitimate power. The combination of low/high levels of coercive power and low/high levels of legitimate power was operationalized through scenarios . Cronbach's α's are excellent and are presented in Table SD = 1.32) or 7.94 USD (SD = 1.69), respectively.Experimental design and procedure were similar to Studies 1 and 2, but four conditions were designed in which the hypothetical tax authority held either low or high levels of coercive power In the following, only hypothesized and/or significant results are reported; however, for completeness, Table low: M = 2.67, SD = 1.21; cphigh: M = 5.50, SD = 1.33; Table Checking the coercive power manipulation with the participants' perceptions of coercive power, the ANOVA showed that low and high levels of coercive power conditions induced respective perceptions  than did participants who experienced low levels of legitimate power . Similar to Study high: M = 4.82, SD = 0.94) rather than low .Likewise, a manipulation check for legitimate power confirmed the manipulation Table . Particiimplicit trust  by 2 (low vs. high levels of legitimate power) ANOVAs with the depending variables implicit and reason-based trust, antagonistic and service climate and enforced compliance and voluntary cooperation were applied. Contrary to expectations, participants in conditions with low/high levels of coercive power and low/high levels of legitimate power reported equal intensity of antagonistic climate increased with coercive power  and decreased with legitimate power  than did low levels of coercive power . Similarly, manipulations of high levels of legitimate power stimulated higher tax honesty intention  than did lower legitimate power  do indicate this significant mediation. All the other indirect effects from coercive and legitimate power are not significant.In a second step, we again used Mediate  majoring in industrial engineering participated in the study. For participation, all students received bonus points for one of their courses. Again, this population primarily was selected because of their naivety regarding the hypotheses and because of their inexperience with insurance organizations.Overall, 102 students  were collected to assess their relative cooperation with the insurance organization and will further be displayed in percentages [(reported amount − 600)/400]. The questionnaire that was used in Studies 1–3 was adapted to the insurance context and was applied to measure insurance fraud intention , implicit trust , reason-based trust , the antagonistic climate , the service climate , enforced compliance , and voluntary cooperation . Analog to the tax context, the participants' perceptions regarding wielded coercive power  and legitimate power  of the insurance organization were adapted for insurance and assessed as manipulation check. Responses were indicated on a seven-point Likert scale ranging from 1  to 7 . Cronbach's α's are excellent and presented in Table In contrast to Studies 1–3, in Study In the following, only hypothesized and/or significant results are reported; for completeness, Table low: M = 3.17, SD = 1.32) than in the high levels of coercive power condition . A weak interaction of coercive and legitimate power on the perceptions of coercive power existed, but as this interaction explains only 6% of the variance and the main effect of coercive power explains 42%, this interaction is negligible. Participants experiencing high levels of legitimate power held perceptions of higher legitimate power  than the participants experiencing low legitimate power  by 2 (low vs. high levels of legitimate power) ANOVAs with the dependent variables implicit and reason-based trust, antagonistic and service climate and enforced compliance and voluntary cooperation were conducted. Similar to the experiments in the tax context, participants in conditions with low/high levels of coercive power and low/high levels of legitimate power reported equal intensity of antagonistic climate increased with coercive power  and decreased with legitimate power  compared to conditions with high levels of coercive power . All other mediation were not significant . The other indirect effects from coercive and legitimate power are not significant.In a second step, we applied again the program Mediate . Furthermore, legitimate power had a positive impact on reason-based trust, the perception of a service climate and voluntary cooperation (Hypothesis 2b). In line with Study Overall, all four studies confirm the hypothesized impact of coercive and legitimate on cognitions when deciding to cooperate with authorities Figure . As expeAs expected, legitimate power (wielded exclusively or in combination with coercive power) increases reason-based trust, the perception of a service climate and the motive to cooperate voluntarily. The relationship of legitimate power and intended cooperative behavior is mediated by reason-based trust.Two unexpected results were found. First, in Study Second, contrary to expectations but in line with earlier findings  might not have that importance for cooperation.The present study was conducted in accordance with the Declaration of Helsinki  and local ethical guidelines for experimentation with human participants at the Faculty of Psychology of the University of Vienna. All participants gave written informed consent prior to the experiment.EH: Research design, conduction of Studies 1–4, analyses of Studies 1–4, drafting of article. BH: Conduction of Study http://www.fwf.ac.at/en/).This research was financed by grant number P24863-G16 from the Austrian Science Fund (FWF) (The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


The aim of this study was to explore the acupuncture effect on the cross-sectional area (CSA) of the median nerve at the wrist in patients with carpal tunnel syndrome (CTS) and, additionally, to identify whether clinical, electrophysiological, and ultrasonographic changes show any association. Forty-five limbs of 27 female patients were randomly divided into two groups (acupuncture and control). All patients used night wrist splint. The patients in the acupuncture group received additional acupuncture therapy. Visual analog scale (VAS), Duruöz Hand Index (DHI), Quick Disabilities of the Arm, Shoulder and Hand (DASH) questionnaire scores, electrophysiologic measurements, and median nerve CSAs were noted before and after the treatment in both groups. VAS, DHI, Quick DASH scores, and electrophysiological measurements were improved in both groups. The median nerve CSA significantly decreased in the acupuncture group, whereas there was no change in the control group. After acupuncture therapy, the patients with CTS might have both clinical and morphological improvement. Carpal tunnel syndrome (CTS) is the most common local entrapment neuropathy resulting from compression of the median nerve as it passes through the carpal tunnel at the wrist. In the general population, CTS prevalence is about 1 to 5% and women are more affected than men . As CTS 2 for median nerve CSA at the wrist level was set with a high sensitivity (% 99) in CTS diagnosis [Musculoskeletal ultrasound (US) has been increasingly used in physiatry practice in the last decade. In CTS, the median nerve can be enlarged with the increase in its cross-sectional area (CSA) in the wrist level due to swelling proximal to the entrapment site. US is an inexpensive and easily accessible imaging method and it can provide morphologic data of the median nerve in patients with CTS. Additionally, CSA of the median nerve was found to be associated with severity of CTS , 7. A cuiagnosis . Furthertivity % 9 in CTS iagnosis . In previagnosis , 12. HowAccordingly, in this study, we aimed to evaluate the effect of acupuncture on median nerve CSA and to explore the correlation between clinical/electromyographic and ultrasonographic changes after acupuncture treatment in patients with CTS.Twenty-seven female patients (45 limbs) with CTS were included in this randomized, controlled study. CTS diagnosis was made using electromyographic studies. The presence of radicular pain, polyneuropathy, radial or ulnar nerve diseases, severe CTS, history of trauma, and previous surgery of the hand and wrist were identified as exclusion criteria. The study protocol was approved by the local ethics committee. All subjects were informed about the study procedure and they provided written informed consent to participate.Patients were randomly separated into two groups: acupuncture and control groups. All subjects were numbered according to the order of admittance to our clinic, and then randomization was performed via a computer program. Both extremities of bilateral CTS patients were included in the same group. All patients used night wrist splint for CTS (set at 0–5 degrees of wrist extension) for 4 weeks, while in the acupuncture group the patients received additional acupuncture treatment. Acupuncture was applied by the same experienced physician; during treatment, the patients were lying on the examination table in the supine position. Nine acupuncture points  which were described in detail previously were chosen for treatment . A 0.25 Age, body mass index (BMI), and disease duration of all patients were recorded. Severity of symptoms, hand function, and electrodiagnostic and US measurements were evaluated before and at the end of treatment. Severity of symptoms was measured by using the visual analog scale (VAS) (0–10 cm). Duruöz Hand Index (DHI) and Quick Disabilities of the Arm, Shoulder and Hand (DASH) scores were used to assess hand functions and disability. DHI consists of 18 questions evaluating activities requiring force, rotational motions, ability, precision, and flexibility of the first three fingers. Each question is scored between 0 and 5 . DHI score was calculated as the sum of the points . Quick DμV), and sensory nerve conduction velocity (SNCV)  were measured. CMAP was enrolled by using skin surface electrodes. The median nerve was stimulated at two different sites (wrist and elbow), and potentials were obtained from abductor pollicis brevis muscle. SNAP was recorded from the wrist via antidromic stimulation of the median nerve at the second finger. According to the electrodiagnostic tests, diagnosis of CTS was classified as mild, moderate, and severe .Electrophysiological tests were performed by using Nihon Cohden Neuropack  machine. Compound muscle action potential (CMAP) , motor nerve conduction velocity (M-NCV) , motor distal latency , sensory nerve action potential (SNAP)  with a 6 to 12 MHz probe. During imaging, patients were seated; the shoulder was kept in neutral rotation, the elbow at 90° of flexion, and the forearm in supination position. CSA of the median nerve was measured at the proximal carpal tunnel during axial imaging. The scaphoid and pisiform bones were identified as bony landmarks for the proximal tunnel, where the CSA was measured .U test or chi-square test was used (where appropriate). Comparison of clinical and electromyographic features and US measurements between pretreatment and posttreatment in each group was performed by using Wilcoxon signed-rank test. Correlations between clinical/electromyographic and US changes were analyzed by using Spearman correlation coefficients. The statistical significance was considered as p < 0.05.For statistical analysis, SPSS  version 22.0 program was used. Data were expressed as mean ± standard deviation. To compare demographic, clinical, electrodiagnostic, and US measurements between the two groups, Mann–Whitney p > 0.05 for all).All patients completed the study. Demographic features of subjects are given in p < 0.05). Additionally, motor distal latency (p = 0.03) and S-NCV (p < 0.001) were increased, and median nerve CSA (p < 0.001) was decreased (p = 0.001) in the acupuncture group, while these parameters did not change significantly in the control group (p > 0.05). Percentage changes of VAS, Quick DASH, DHI scores, S-NCV, and CSA of the median nerve were higher in the acupuncture group compared to the control group  (p > 0.05). The percent changes in the severity of CTS did not differ in both groups.Clinical characteristics and electromyographic and US measurements are summarized in for all) . ChangesThe aim of this study was to explore whether acupuncture treatment had any additional effect on median nerve morphology in patients with CTS via US imaging. The results of this study showed that CSA of the median nerve decreased after acupuncture treatment. Additionally, symptom severity, hand functions, and electromyographic measurements improved in both groups. However, the improvements in VAS, DHI, Quick DASH, and S-NCV were significantly higher in the acupuncture group. The change in median nerve CSA did not correlate with clinical and electromyographic changes. While previous studies investigated the effect of acupuncture on clinical symptoms and electromyographic studies, to the best of our knowledge, its effect on median nerve morphology was not investigated before.The mechanism of acupuncture effect on CTS has not been identified clearly yet. However, in recent studies, with the help of magnetic resonance imaging, it has been shown that acupuncture treatment might alter brain activity or regulate limbic system activity in patients with CTS –18. Addi2 may indicate the presence of CTS [Ultrasound measurement of median nerve CSA was used for diagnosis and assessment of response to therapy in patients with CTS –8, 10. Ce of CTS , 7. In oThe relatively small sample size and the lack of long-term monitoring of patients are major limitations of our study. Moreover, placebo effects of acupuncture cannot be evaluated. Sham acupuncture effect for CTS is debated. The effect of sham acupuncture in CTS in previous studies is conflicting. In a randomized controlled study, it was reported that acupuncture was not more effective than sham acupuncture when applied together with night splints . In anotTo summarize, in light of the results of the current study, we showed that acupuncture improves clinical and electrophysiological findings of CTS and also decreases CSA of the median nerve at the wrist. Our results need to be confirmed in future studies with larger sample, long-term monitoring, and placebo controlled studies. Finally, US seems to be a practical imaging tool for diagnosis and monitoring in these patients.


Mycobacterium tuberculosis, changes in gene expression are highly variable and involve many genes, so traditional single-gene screening of M. tuberculosis targets has been unable to meet the needs of clinical diagnosis. In this study, using the National Center for Biotechnology Information (NCBI) GEO Datasets, whole blood gene expression profile data were obtained in patients with active pulmonary tuberculosis. Linear model-experience Bayesian statistics using the Limma package in R combined with t-tests were applied for nonspecific filtration of the expression profile data, and the differentially expressed human genes were determined. Using DAVID and KEGG, the functional analysis of differentially expressed genes  and the analysis of signaling pathways were performed. Based on the differentially expressed gene, the transcriptional regulatory element databases (TRED) were integrated to construct the M. tuberculosis pathogenic gene regulatory network, and the correlation of the network genes with disease was analyzed with the DAVID online annotation tool. It was predicted that IL-6, JUN, and TP53, along with transcription factors SRC, TNF, and MAPK14, could regulate the immune response, with their function being extracellular region activity and protein binding during infection with M. tuberculosis.Tuberculosis (TB) is one of the deadliest infectious diseases worldwide. In  Mycobacterium tuberculosis, and 5–10% of these patients have active disease  [M. tuberculosis and with the original CEL documents available were selected.The gene chip array data were retrieved from GEO Datasets of the National Center for Biotechnology Information  database using keywords “tuberculosis” and “We chose gene expression profile of GSE16250 from GEO database, which was a public and freely available database. GSE20050 was based on agilent GPL1352 platform. The GSE52819 dataset included 12 samples, containing 6 TB samples and 6 healthy samples. We also downloaded the Series Matrix File of GSE54992 from GEO database .M. tuberculosis was compared by SAM, and the genes with fold change > 2.0 or <−2.0 and p value < 0.05 were selected as differentially expressed genes. The differentially expressed genes were further screened by a Venn diagram; considering the differences between the chip platforms, the genes that overlapped with three or more than three platforms were selected.The Affymetrix Expression Console software tool was applied for the background correction of the collected chip data and the conversion of probe fluorescence values into gene expression values . The Affhttp://rulai.cshl.edu/TRED [Transcription factor and target gene screening analysis was carried out at the TRED website edu/TRED .Eighteen transcription factors (TF) and 887 target genes were predicted to a combined total of 887 TF-to-target pairs. The relation obtained from the analysis on the differential coexpression was mapped to the human transcription factors and target gene pairs to obtain transcription regulation pairs. Finally, Cytoscape software was used for plotting .This analysis was carried out using DAVID database at website. 1075 genes were included in the analysis, and the human genome was used as the source of background genes .χ2 test were used to calculate the significance level and misdiagnosis rate of each GO, and the p value was calibrated with the misdiagnosis rate, thereby screening out the GO significance reflected by the differentially expressed genes (p < 0.05). Using the European Bioinformatics Institute (EBI) database, the experimental results were manually analyzed.Based on the NCBI Gene Ontology database, the GO annotation of these genes was performed to obtain all GO terms which genes were involved . Fisher'http://www.genome.jp/kegg/ in September 2016 [All pathways used in this study were downloaded from KEGG databases. In total, 78 relevant KEGG pathways were downloaded from ber 2016 .M. tuberculosis based on Affymetrix gene chips with the original CEL document were selected, and 4 groups of gene chips were determined, as shown in All the chip data in this study were taken from GEO Datasets of the National Center for Biotechnology Information (NCBI) database. The gene expression data of humans infected with These 4 groups of chips included those whose chip number of chips was 6, 7, 12, and 21, and each chip group was divided into two subgroups, a healthy subgroup and TB patient subgroup.M. tuberculosis (TB patient group) was compared by the SAM method to determine differentially expressed genes. The number of differentially expressed genes of each platforms was 5569, 2442, 616, and 578, respectively. These genes were cross-screened again, and the results showed that there were 130 differentially expressed genes on the 4 platforms, 224, 63, 504, and 154 on three platforms, respectively, with a total of 1075 genes  and those infected with 75 genes .http://rulai.cshl.edu/TRED), transcription factors interacting among the 1075 expressed genes were predicted and analyzed, and 887 genes were found to correspond to 18 transcription factors  database  people worldwide are infected with M. tuberculosis, but only 10% will develop active tuberculosis and the remaining 90% will make the disease to curb due to the body's own well-developed immune system [M. tuberculosis infection will provide a theoretical basis for the progress and prevention of active tuberculosis.Tuberculosis caused by the infection of e system . Our pree system . The dise system . To undeM. tuberculosis pathogenic genes were found. The TRED database was used to predict and analyze these 1075 genes, and as a result, 18 transcription factors and 887 genes were found, and an M. tuberculosis pathogenic gene transcription regulation network diagram was constructed  regulatory elements, RELA plays a key role [M. tuberculosis with inactivated RELA cannot survive for a long time in an environment with the abovementioned types of pressure [M. tuberculosis active propagules; persisters isolated from the lungs show extremely close manifestations to those under conditions of pressure, especially lack of nutrition, which is completely consistent with the starting condition of RELA. Another study showed that RELA is closely related to the bacterial morphology and colony formation caused by changes in the cell wall of mycobacterium, which is important for the formation of M. tuberculosis L-forms, namely, cell wall-deficient M. tuberculosis. RELA is thus a good target against persistent tuberculosis infection [key role –29. Studpressure . There anfection , 32.FOS gene is a member of FOS family, along with JUN family members and other activated transcription factor protein family members including activator protein 1 (AP-1). AP-1 is an important transcription regulation factor in the nucleus, playing an important role in the transduction process of many signals in the body, and is the intersection of a series of nuclear cellular signal transduction pathways . FOS canThe statistics of network nodes is shown in M. tuberculosis in a nonproliferative state in the body of the host, thereby resulting in a delayed course of the disease and a recurrent persistent infection.In summary, 1075 genes were screened out through GEO database and a transcription factor regulation network diagram was constructed. Through the statistical analysis of the transcription factors and network nodes of their regulatory network diagram, we speculate that IL-6, JUN, and TP53, along with transcription factors SRC, TNF, and MAPK14, could regulate the human immune function, evading the attack on the host immune system and ultimately leading to the existence of