Reliable and objective assessment of psychomotor skills in physiotherapy students’ education is essential for direct feedback and skill improvement. The aim of this study is to determine the interrater reliability in the assessment process of physiotherapy students and to analyse the assessment behaviour of the examiners.Physiotherapy teachers from two different schools assessed students from two different schools performing proprioceptive neuromuscular facilitation (PNF) patterns. An evaluation sheet with a 6-point rating scale and 20 evaluation criteria including an overall rating was used for assessment. The interrater reliability was determined calculating an intraclass-correlation coefficient (ICC) and Krippendorff’s alpha. The assessment behaviour of the examiners was further analysed calculating the location parameters and showing the item response distribution over item in form of a Likert plot.The ICC estimates were mostly below 0.4, indicating poor interrater reliability. This was confirmed by Krippendorff’s alpha. The examiners showed a certain central tendency and intergroup bias.The interrater reliability in this assessment format was rather low. No difference between the two physiotherapy schools concerning the interrater reliability could be identified. Despite certain limitations of this study, there is a definite need for improvement of the assessment process in physiotherapy education to provide the students with reliable and objective feedback and ensure a certain level of professional competence in the students.The study was approved by the ethics committee of the Medical Faculty RWTH Aachen University (EK 340/16).The online version contains supplementary material available at 10.1186/s12909-022-03231-y. Reliability and validity in assessing clinical or practical skills is a continuous challenge in the education of health care professionals such as physiotherapists .The term “clinical skills” is not unanimously defined. The definitions include different aspects such as practical procedures and communication skills among others . They arAlmost every discipline in the health care sector has to face the challenge of teaching psychomotor skills . Health The same importance of psychomotor skills applies for the physiotherapy profession, where the taught skills are complex. One skill consists of a variety of single components, which must be executed simultaneously to achieve the required effect in the patient’s treatment . To impaIn the last decades, there have been ambitions in the community of physiotherapists in Germany to standardize the assessment process of these skills, for example, by using the mini-Clinical Evaluation Exercise (mini-CEX) assessment format. The mini-CEX has already proven to be a valid and reliable tool for assessing medical students .A key issue in the assessment process is the objectivity of examiners and the risk of examiner bias . The obsThe training as a physiotherapist in Germany lasts 3 years and is completed with a state examination. Psychomotor skills are essential to the education and an integral part of the mandatory coursework . LearninTo give insights into the current status of the quality of this feedback, the aim of this study is to assess the interrater reliability of the evaluation of psychomotor skills in the education of physiotherapy students based on an assessment tool originally created for peer evaluation (mini-PEX) in lieu of a mini-CEX evaluation. In addition, we took a brief look at the examiners’ assessment behaviour.This study is part of the “Media Didactics Meets Wearable Computing (MediWeCo)” project (project number FKZ 01PD15013) founded by the German Federal Ministry of Education and Research (BMBF) through the program “Digital media in vocational education” supported by the European Union Social Fund (ESF). MediWeCo aims to develop digital learning alternatives to support and improve the process of teaching, learning, and assessing psychomotor skills in physiotherapy.We conducted a study to determine the interrater reliability of the evaluation process in physiotherapy education and to analyse the examiners’ assessment behaviour. Specifically, physiotherapy teachers evaluated physiotherapy students performing proprioceptive neuromuscular facilitation (PNF) patterns. PNF is a widely used treatment by physiotherapists and forms part of the curriculum in many countries . After iThe study was approved by the ethics committee of the Medical Faculty RWTH Aachen University (EK 340/16). A total of three examination dates was scheduled, two at the “Schule für Physiotherapie Uniklinik RWTH Aachen” (AC) and one at the “Grone Berufsfachschule für Physiotherapie Hamburg” (HH).A total of 47 students participated in the trial, 21 and 26 from HH and AC respectively. The physiotherapy students were in their first year of training and were instructed in PNF. All students gave their written consent to participate in this study. The AC students had two examination dates with a total of 41 exams as AC students were absent on the day of the examination in 11 cases. The HH students had one examination date with 21 examinations. Each examination was video recorded with a total of 62 videos.Six physiotherapy teachers participated in this study, three from HH and three from AC. The six physiotherapy teachers were anonymised and numbered consecutively from 1 to 6, with AC teachers having numbers 1 to 3 and HH teachers having numbers 4 to 6.Three PNF leg patterns were selected: 1) extension-abduction-internal rotation with knee extension, 2) extension-adduction-external rotation with knee extension, 3) flexion-adduction-external rotation with knee flexion.The first two patterns were performed by the AC students, the third by the HH students. We chose different patterns for the two schools to avoid competition and direct comparison between the two schools in hopes of minimizing intergroup bias. All three patterns were executed using the technique Rhythmic Initiation consisting of four phases: Passive, Passive-Active, Resistive, and Active phase. Each phase of this technique was repeated 10 times.The video recordings of the students’ performance were evaluated afterwards by four examiners out of the group of six examiners. The four examiners were chosen randomly for every single performance, two out of the three from HH and two out of the three from AC. Because of the workload it was not possible for every examiner to evaluate every performance. The order in which the video recordings were evaluated was not defined. For the evaluation, the examiners used an evaluation sheet developed within the MediWeCo project, consisting of 20 evaluation criteria. The 20 criteria consist of 19 single criteria and one overall rating criterion. The evaluation sheet was originally designed as a peer evaluation tool, a so called mini-Peer Evaluation Exercise (mini-PEX) and used at the “Schule für Physiotherapie Uniklinik RWTH Aachen” (AC). We used it as a mini-CEX evaluation tool. A publication concerning the validity and reliability of this tool is pending. It was not the aim of this study to assess the validity of the evaluation tool. Each evaluation criterion was rated on a six-point Likert scale from 1 to 6, with 6 corresponding to a completely fulfilled criterion. The 20 evaluation criteria are: Bench Height, Treatment Area, Patient Position, Verbal Communication, Explanation, Rhythm, Passive, Active-Assistive, Resistive, Active, End Position, Diagonal, Movement Components, Timing, Body Position, Body Mechanics, Lumbrical Grip, Stimulus, Resistance, Overall Rating.To determine the interrater reliability between the examiners, the intraclass correlation coefficient (ICC) was calculated. The ICC is a standard tool to determine the interrater reliability of more than two raters on the basis of interval scaled data. The six-point rating scale used in this study is interval-scaled.To compare the single ratings from each examiner, ICC estimates and their 95% confidence intervals were calculated using IBM SPSS Statistics for Macintosh, Version 25.0. , based on a single-rating, absolute agreement, one-way random effects model  –15. To cFor the single ratings, 15 ICC estimates were calculated, one between examiner 1 and 2, 1 and 3, and so on until examiner 5 and 6. Each ICC was calculated separately for every evaluation criterion. Table To verify the ICC estimates, we additionally calculated Krippendorff’s alpha. Krippendorff’s alpha was interpreted by the guidelines suggested by Krippendorff, with α ≥ 0.800 indicating good interrater reliability, 0.800 > α ≥ 0.667 only allowing for tentative conclusions, and α <  0.667 suggesting poor interrater reliability .To analyse the examiners’ assessment behaviour and to illustrate the distribution of ratings from AC and HH, we calculated the location parameters for the Overall Rating for the two groups (AC and HH) of examiners and students and visualized the data in form of box plots Fig. . AdditioOverall Rating) is 0.559   in the category Overall Rating signifying an excellent interrater agreement between examiner 5 and 6. The ICC 2,4 Overall Rating estimate has the lowest value of 0.011 in the category Overall rating (CI -0.347 – 0.369) signifying a poor interrater agreement.The ICC estimate comparing the mean rating of the AC examiners and the mean rating of the HH examiners for the criterion Overall Rating (ICC AC vs HH 99) Fig. . This siComparing the ratings of the HH examiners and AC examiners, no systematic difference in the ICC estimates was obvious Fig. . The ICCIn the group of AC examiners  44 (73.3%) of the altogether 60 ICC estimates fall into the category poor, 11 (18.3%) into the category fair, 4 (6.7%) into the category good, and 1 (1.7%) into the category excellent. In the group of HH examiners  37 (61.7%) of the altogether 60 ICC estimates fall into the category poor, 8 (13.3%) into the category fair, 7 (11.7%) into the category good, and 8 (13.3%) into the category excellent. Comparing AC and HH examiners (ICC AC vs HH) 11 (55%) of the altogether 20 ICC estimates fall into the category poor, 4 (20%) into the category fair, 3 (15%) into the category good, and 2 (10%) into the category excellent.Overall, the ICC estimates vary between poor and excellent depending on which criterion is rated and by whom it is rated, but with the estimates indicating poor interrater reliability being predominant. An additional table shows this in more detail  signify a poor interrater reliability for 11 evaluation criteria , a fair interrater reliability for 4 evaluation criteria , a good interrater reliability for 3 evaluation criteria , and an excellent interrater reliability for 2 evaluation criteria  , whereas the median for AC students rated by HH examiners is 4 (box plot 1). The lower quartile for HH students rated by HH examiners (box plot 2) is also higher than for AC students rated by HH examiners (box plot 1). However, the minimum for HH students rated by HH examiners (box plot 2) is one point lower than for AC students rated by HH examiners (box plot 1).The median for AC students rated by AC examiners and HH students rated by AC examiners is the same, but the lower quartile and the minimum for AC students rated by AC examiners is higher than for HH students rated by AC examiners (box plot 4 and 5).The interquartile range varies between 1  and 2 , meaning that 50% of the ratings lie in the range between 4 and 5, and 3 and 5 respectively. In box plot 4 and 6 the range between lower quartile and median is equal zero, meaning the 25 and 50 percentiles are equal.Minimum and maximum are equal 1 and 6, respectively, in box plots 2, 3, 5 and 6. Looking at the ratings of AC students given by HH examiners and by AC examiners minimum and maximum are equal 2 and 6, respectively (box plots 1 and 4). The worst rating given to AC students by AC examiners and by HH examiners is one point higher than the worst rating given to HH students by AC examiners and by HH examiners . The best rating given is the same no matter which students are rated or by whom they are rated.The Likert plot  lie beneath the mark of 0.4  is the criterion Explanation. The examiners’ expectation of the students seems to be very similar concerning this evaluation criterion, whereas the concept of what is a good or bad performance seems to differ strongly for the criteria Passive, Timing, and Resistive. All ICC estimates for these three evaluation criteria fall into the category poor. One could assume, that verbal performance like the criterion Explanation is easier to assess objectively and reliably than motoric performance like the criteria Passive, Timing, and Resistive.In the end, the rating that a physiotherapy student receives for his/her performance seems to be heavily dependent on which examiner is rating the student’s performance. But a positive aspect is that the interrater reliability seems not to be dependent on which school or location the examiner is from. Yet, our results suggest that the interrater reliability is in high need of improvement.The boxplots comparing the scores for the Overall Rating from HH and AC examiners show a certain intergroup bias or favouritism Fig. . HH examThe better ratings of in-group students may also be caused by the difference in expectations and style of teaching. HH students are taught by HH examiners and know which performance they expect and the same applies for AC students. Students may be able to fulfil the particular expectations of their own examiners more completely than the unknown expectations of external examiners.Another aspect seen in the box plots Fig.  concerniAn explanation for this rating behaviour could be the homogeneity of the group of participants. The students are all in their first year of training and on the same skill level, therefore it could be that most of their performances truly lie in the middle of the rating scale. The clustering of ratings in the middle does not necessarily have to be caused by central tendency bias.When interpreting ICC estimates, some pitfalls have to be considered. There are some limitations to the validity of the ICC estimates in this study, one being the relatively small group of participants of 47 physiotherapy students. But as a thumb rule it is suggested that 30 participants are deemed sufficient in reliability studies .In addition, not every physiotherapy student was evaluated by all 6 physiotherapy teachers. One can discuss, that if every teacher would have assessed every student’s performance, this would have led to more data and a potentially more significant ICC estimate, as an ICC estimate comparing ratings of all 6 teachers could have been calculated using a “consistency”  or respeThe ICC model we used was a one-way random effects, absolute agreement, single rater and multiple rater model, which usually produces lower ICC estimates than the consistency model mentioned above . This haIn addition, decisions about the professional competence, for example, which applicant gets the job, are usually made based on the absolute values of the rating and do not consider differences in the mean values of different raters. This is another reason why the absolute agreement model seems to be more appropriate in this context.As we used a single and multiple rater model, with the multiple rater model based on mean values from the HH examiners and mean values from the AC examiners, one also has to be aware of the fact that a multiple rater model generates higher ICC estimates than a single rater one . This coAnother factor leading to poor interrater reliability could be the lack of variance within the group of physiotherapy students. If the group of participants is too homogeneous in their skills, the ICC estimates underrate the true interrater reliability . This coThe existing literature suggests that the training or briefing of the participating raters is one possible way to influence the outcome of interrater reliability. Some sources suggest that rater training improves interrater reliability –25, wherUsing different assessment formats could be an additional way to improve interrater reliability and make the rating process more objective. The Objective Structured Clinical Examination (OSCE) was introduced in the 1970s by Harden  and is wAside from the OSCE format, standardized assessment tools like the “Assessment of Physiotherapy Practice” tool (APP) could be another possibility to increase reliability, but they are mostly used for an assessment over a longer period of time in clinical settings , 35 .Finally, smart wearables with movement sensors could support the assessment process of psychomotor skills and increase its objectivity not only in the field of physiotherapy. Some sources showed the successful application of smart wearables in the training and evaluation of hand hygiene , knee arDespite all restrictions concerning the ICC estimates, this study still shows a mostly poor interrater reliability in the assessment process in physiotherapy education. The rating a student receives is heavily dependent on the examiner the student is rated by. No significant difference in the interrater reliability between the two schools could be detected. These findings are confirmed by the two methods ICC and Krippendorff’s alpha. Besides the examiners’ assessment behaviour shows a certain in-group favouritism and central tendency bias when looking at the rating distribution. Overall, the physiotherapy students do not receive an assessment as reliable as they need to improve their skills properly. Collaborative efforts in the physiotherapy community concerning the assessment process should be made to receive more reliable measurements of physiotherapy students’ performance. Contributions and input from leading teaching institutions are necessary.A form of standardized assessment to improve reliability, international or at least national, would be desirable. Different assessment formats as well as the use of automated, digital assessment tools could be a possibility to improve reliability in the assessment process of psychomotor skills in physiotherapy and other health care professions and should be the subject of further study.Additional file 1. ICC estimates in categories.Additional file 2. Krippendorff’s alpha estimates.Additional file 3. Likert plot.Additional file 4: Addendum A. ICC estimates by evaluation criterion.


Few studies have reported whether a biopsy in emergency gastroscopy (EG) increased the risk of rebleeding in patients with Forrest I acute nonvariceal upper gastrointestinal bleeding (ANVUGIB) combined with suspected malignant gastric ulcer (SMGU). This study aims to conduct a multicenter retrospective cohort study using propensity score matching to verify whether a biopsy in EG increases the risk of rebleeding in patients diagnosed with Forrest I ANVUGIB combined with SMGU.Using the data for propensity-matched patients, logistic regression models were fitted using rebleeding as the dependent variable. Survival time was defined as the length of time the patient experienced from visiting the emergency department to rebleeding. We used the Kaplan–Meier (KM) method to analyze the 30-day survival of the patients with and without a biopsy after matching, and the log-rank test was performed to examine the differences in survival.With the use of propensity score matching, 308 patients who underwent a biopsy in EG were matched with 308 patients who did not. In the five logistic regression models, there were no significant group differences in the risk of rebleeding in patients with Forrest I ANVUGIB combined with SMGU between the biopsy and no-biopsy groups. The probability of survival was not significantly different between the no-biopsy and biopsy groups.In this multicenter, retrospective propensity score matching cohort study, compared with patients without a biopsy, patients with a biopsy during EG had no increased risk of rebleeding, and there was no significant difference in the rate of rebleeding. Acute nonvariceal upper gastrointestinal bleeding (ANVUGIB) is one of the most common acute critical diseases in clinical practice. The morbidity of ANVUGIB in Europe ranged from 25/100,000 to 35/100,000 in 2000, and the morbidity of ANVUGIB in the United States was 60.6/100,000 in 2009, among which the morbidity of ANVUGIB caused by peptic ulcers was 32.1/100,000 , 1211, 1Rebleeding was defined as one or more signs of ongoing bleeding, including hematemesis, melena, hematochezia, vital sign instability and a continuous drop in hemoglobin after the initial resuscitation or initial endoscopic therapy of the patient, which required repeated EG, angiographic embolization or operation to stop the bleeding , 12.The primary outcome of the study was the risk of hospital rebleeding the secondary outcome was hospitalization costs. These patients were followed throughout the hospital course until in-hospital rebleeding episodes.PASS 15  was used to calculate the sample size for the cohort study. operation procedure: Proportion → Two Independent Proportions → Test  → Tests for Two Proportions (Ratios). According to guidelines, the rebleeding rate of Forrest I ANVUGIB is 55%, and we hypothesized that a biopsy has a low risk effect (RR = 1.2) at increasing rebleeding compared with no biopsy , 13. To We used logistic and linear regression analyses after propensity score matching to control for confounding factors in this real-world study.Given the differences in the baseline characteristics between eligible participants in the two groups Table , propens2 or Fisher’s exact tests. Using the data for the propensity-matched patients, logistic regression models were fitted using rebleeding as the dependent variable. Using the data for the propensity-matched patients, multivariate linear regression models were fitted using hospitalization costs as the dependent variable.Group differences were evaluated with Mann–Whitney U-test, Student’s t test, and χA two-tailed P value < 0.05 was considered significant. All analyses were conducted using SPSS 23.0 .To test the robustness of the main results, several additional analyses were conducted. First, multiple imputation using multivariate normal distribution was performed to evaluate the potential influence of missing data. Second, using the data for all the patients with Forrest I ANVUGIB combined with SMGU before matching, logistic regression models were fitted using rebleeding as the dependent variable. Third, using the data for all the patients with Forrest I ANVUGIB combined with SMGU before matching, multivariate linear regression models were fitted using hospitalization costs as the dependent variable. Fourth, subgroup analysis with the data before and after matching was also conducted by stratifying patients into the non-taking antithrombotic group, aspirin-alone group and clopidogrel-alone group.Survival time was defined as the length of time the patient experienced from visiting the emergency department to rebleeding. We used the Kaplan–Meier (KM) method to analyze the 30-day survival without rebleeding of the patients with and without a biopsy after matching, and the log-rank test was performed to examine the differences in survival.Figure A total of 1010 patients constituted the study cohort, of whom 401 (39.7%) underwent a biopsy and 609 (60.3%) did not. Before propensity score matching, there were differences between the two groups in several of the baseline variables Table . With thIn the cohort before and after matching, there were no significant differences in the risk of rebleeding in patients with Forrest I ANVUGIB combined with SMGU during EG between a biopsy and no biopsy Table .Table 3RTable In the cohort before matching, there was no significant difference in the hospitalization costs of patients with Forrest I ANVUGIB with SMGU during EG between biopsy and no biopsy. In the cohort after matching, the hospitalization costs of patients undergoing a biopsy with Forrest I ANVUGIB combined with SMGU during EG were significantly lower than those of patients who did not undergo a biopsy Table .Table 5HTable The multiple imputation presented a familiar consequence: there were no significant differences in the risk of rebleeding in patients with Forrest I ANVUGIB combined with SMGU during EG between biopsy and no biopsy Table ; a biopsWe also performed statistical analysis on all patients before matching and obtained similar results. There were no significant differences in the risk of rebleeding in patients with Forrest I ANVUGIB combined with SMGU during EG between biopsy and no biopsy groups Table . A biopsSubgroup analysis with the data before and after matching was also conducted by stratifying patients into the non-taking antithrombotic group, aspirin-alone group and clopidogrel-alone group. In all subgroup analyses, there were no significant differences in the rate of rebleeding between patients who underwent a biopsy in EG and those who did not Table .Table 7SFigure To our knowledge, this is the first multicenter retrospective cohort study that focused on biopsies during EG in patients with Forrest I ANUGIB combined with SMGU using a propensity-matched approach. The present results demonstrated that there were no significant differences in the risk of rebleeding in patients with Forrest I ANVUGIB combined with SMGU during EG between biopsy and no-biopsy groups. The hospitalization costs of patients undergoing a biopsy with Forrest I ANVUGIB combined with SMGU during EG were significantly lower than those of patients who did not undergo a biopsy.ANVUGIB accounts for 80%–90% of UGIB, with peptic ulcers, acute gastric mucosal injury and upper gastrointestinal tumors as the most common causes . In receTo answer these most difficult questions that clinicians encounter in EG, we selected patients with Forrest I (the level with the highest risk of rebleeding) ANVUGIB to constitute the cohort. Active bleeding in the lesion can be seen during gastroscopy in patients with Forrest I ANVUGIB, and the patient is in a critical situation and must undergo endoscopic hemostatic therapy , 21. HowOur study showed that in univariate analysis after matching and multivariate analysis before and after matching, the hospitalization costs of patients undergoing biopsy with Forrest I ANVUGIB combined with SMGU during EG were significantly lower than those of patients who did not undergo a biopsy. According to the guidelines and consensus recommendations, benign and malignant lesions determine the difference in the treatment paths of patients with ANVUGIB . Timely According to the guidelines and consensus, in routine gastroscopy, a biopsy is a low-risk operation, aspirin- or clopidogrel-alone can be continued before a biopsy, and there is no significant difference in the rate of rebleeding with the withdrawal of antithrombotic , 13, 17.Some limitations of our analysis should be considered. First, this was a nonrandomized, observational study and hence suffers from potential selection and ascertainment bias, despite robust propensity score matching. Second, we did not compare the risk of death between the biopsy and the no-biopsy groups because ANVUGIB has a mortality rate of 1.1% to 1.7%, and our current sample size does not support a comparison of the risk of death. In our cohort, before matching, there were 6 deaths (6/401 1.5%) in the biopsy group and 8 deaths (8/609 1.3%) in the no-biopsy group; after matching, there were 3 deaths (3/308 1.0%) in the biopsy group and 5 deaths (5/308 1.6%) in the no-biopsy group. In the future, we will continue to include data from more centers to have a sufficient sample size to analyze the risk of death.In this multicenter, retrospective propensity score matching cohort study of patients with Forrest I ANVUGIB combined with SMGU, compared with patients without a biopsy, patients with a biopsy during EG had no increased risk of rebleeding, and there was no significant difference in the rate of rebleeding. Patients with a biopsy had significantly lower hospitalization costs than those without a biopsy.


From a reproductive justice framework, we aimed to investigate how a possible association between hormonal contraceptive (HC) and antidepressants use (as a proxy for depression) is distributed across intersectional strata in the population. We aimed to visualise how intersecting power dynamics may operate in combination with HC use to increase or decrease subsequent use of antidepressants. Our main hypothesis was that the previously observed association between HC and antidepressants use would vary between strata, being more pronounced in more oppressed intersectional contexts. For this purpose, we applied an intersectional multilevel analysis of individual heterogeneity and discriminatory accuracy approach.Observational prospective cohort study using record linkage of national Swedish registers.The population of Sweden.All 915 954 women aged 12–30 residing in Sweden 2010, without a recent pregnancy and alive during the individual 1-year follow-up.Use of any antidepressant, meaning being dispensed at least one antidepressant (ATC: N06A) during follow-up.Previously mentally healthy HC users had an OR of 1.79 for use of antidepressants compared with non-users, whereas this number was 1.28 for women with previous mental health issues. The highest antidepressant use were uniformly found in strata with previous mental health issues, with highest usage in women aged 24–30 with no immigrant background, low income and HC use (51.4%). The largest difference in antidepressant use between HC users and non-users was found in teenagers, and in adult women of immigrant background with low income. Of the total individual variance in the latent propensity of using antidepressant 9.01%  and 8.16%  was found at the intersectional stratum level.Our study suggests teenagers and women with immigrant background and low income could be more sensitive to mood effects of HC, a heterogeneity important to consider moving forward. Entire Swedish population of women aged 12–30 included.Pharmacy dispensing automatically linked to individual personal identification number in Sweden through the Swedish Prescribed Drug Register and thus very reliable.Intersectional multilevel analysis of individual heterogeneity and discriminatory accuracy is a fruitful way of epidemiologically investigating heterogeneity within a population while considering individual conditions determined by societal power dimensions such as class, gender and race.Antidepressant dispensing is not a perfect proxy for depression.Registers cannot measure actual use of any medication.In recent years, attention in the medical community has increasingly been drawn towards depression and other adverse effects on mood related to use of hormonal contraception (HC).A fruitful way of epidemiologically investigating heterogeneity within a population while considering individual conditions determined by societal power dimensions such as class, gender and race has been developed through intersectional theory in recent years.To operationalise an intersectional mapping of heterogeneity in use of antidepressants in relation to HC on a population level, we used a multilevel analysis of individual heterogeneity and discriminatory accuracy (MAIHDA).We conceptualise the intersectional strata as social contexts rather than static individual traits, thereby visualising how intersecting power dynamics can act in combination with HC to predispose for depressive mood. Our main hypothesis was that the previously observed association between HC and use of antidepressants would vary between strata and that this association would be more pronounced in more oppressed intersectional contexts. We investigate this hypothesis on the whole population of women susceptible to HC use in Sweden.After allowance from the Swedish Ethical Authority and the data safety committees from Statistics Sweden and the Swedish National Board of Health and Welfare, we obtained a database created by record linkage of several nationwide registers administered by Statistics Sweden  and the Swedish National Board of Health and Welfare  and the Cause of Death Register). The Swedish authorities linked the registries using a unique personal identification number, but the database was anonymised before delivering it to us.We defined an initial cohort containing all 1 064 171 women aged 12–30 years residing in Sweden 1 January 2010 and obtained individual level data on medication use from SPDR, which contain all dispensed drug prescriptions at Swedish pharmacies since 2006.Every woman was assigned an individual baseline date, defined by the first dispensed prescription of an HC drug between 1 January 2010 and 31 December 2014 after 12 years of age, and was then followed for 1 year after her individual baseline date. A woman obtaining her first prescription on 1 September 2013 was therefore followed to 1 September 2014. For non-users of HC, the baseline date could not be based on a HC-prescription and was therefore assigned, to 1 July 2012 for all adults, but later for some of the younger girls turning 12 during our period of investigation. This means all non-users had been true non-users for at least 1.5 years before their follow-up started (1 January 2010 to 1 July 2012) but also continued to be non-users all the way to 31 December 2014. From the individual baseline date, the women were followed for 1 year to find out if a prescription of an antidepressant was dispensed. Data were also collected on psychiatric disorders and psychotropic drug use in the past 3 years (see Assessment of variables). After excluding women with incomplete follow-up time due to death, emigration, missing information on country of birth, and pregnancies 1 year before and after the baseline as well as, the final database consisted of 915 952 women. This database was divided into two cohorts according to the presence or absence of previous mental health issues, see Users of HC were defined as any women who, according the SPDR, filled a prescription of HC  classification system codes G02B, G03AA-C) between 1 January 2010 and 31 December 2014, while non-users did not have a prescription filled during the same period. Emergency contraception (G03AD) that are mainly bought over the counter in Sweden was excluded. The majority of HC prescriptions are acquired via midwifes in Sweden , whom can only prescribe HC for contraceptive purposes. Physicians, most often gynaecologists, can also can prescribe HC for other purposes such as in response to bleeding disturbances or endometriosis. Since these indications could confound our results, we excluded women with physician-issued prescriptions, see Antidepressant use, the outcome of our study, was defined, according to the SPDR, as being dispensed at least one prescription of antidepressants (ATC: N06A) during the individual 1-year follow-up.Previous mental health issues were defined as having any psychiatric disorder diagnosed at a hospital (ICD: F00-F99) or a dispensed prescription of a psychotropic drug  in the past 3 years.Pregnancies 1 year previous to baseline and during follow-up were identified according to the 2019 version of the Nordic Diagnosis-Related Group classification (NordDRG), Major Diagnostic Categories codes M14 for pregnancy, delivery and postpartum care.We used family level data on income as of 31 December 2010 from Statistics Sweden’s LISA. Individualised disposable family income was calculated by dividing the total disposable income of the family by the number of family members, taking into account the different consumption weights of adults and children determined by Statistics Sweden. Thereafter, we created three categories  of income using tertile cut-offs based on the total Swedish population aged 18–80 years. We considered the high-income category as the reference in the comparisons.We defined immigrant status at the family level as no family member >18 years of age born in Sweden, since understanding of and access to institutions such as healthcare differ depending on social position such as it is constructed by the power dimensions of race/immigration, as well as the experience of xenophobia. This variable should therefore be considered as an effort to capture a social position affecting possibilities and life trajectories rather than an essentialist view of otherness. We categorised age at the individual baseline into the following groups: 12–17, 18–23 and 24–30 years to capture age specific conditions of adolescents, young adults and adult women.Within each cohort stratified by previous mental health issues, we generated 36 intersectional strata by combining three categories of age, three categories of income, two categories of immigrant background and two categories of HC use. Mental health issues can be considered as a valid category of intersectional investigation in a society that considers an able body and mind vital, in other words relating to the power dimension of able-bodiedness.We performed an intersectional MAIHDA with individual women at the first level of analysis and the 36 intersectional strata at the second level, stratified by previous mental health issues see . The use10.1136/bmjopen-2021-049553.supp1Supplementary data10.1136/bmjopen-2021-049553.supp2Supplementary data10.1136/bmjopen-2021-049553.supp3Supplementary data10.1136/bmjopen-2021-049553.supp4Supplementary dataThe first model included only an intercept and a random effect for the intersectional strata with no covariates. In this model 1, we first (i) performed a simple analysis of components of variance and calculated the variance partition coefficient (VPC). That is, the share (expressed as a percentage) of the total individual variance in the latent propensity of antidepressant use, that is, at the intersectional strata level. In this simple model, the VPC correspond with the intraclass correlation coefficient which informs on the clustering of antidepressant use within intersectional strata. The VPC values extend from 0% to 100%. Second, (ii) we calculate the stratum-specific absolute usage of antidepressants and their 95% credible intervals (CI) by transformation of the information from the logistic regression to the probability scale. We used this information to map the user heterogeneity across the intersectional strata. Then, (iii) using these stratum-specific predictions, we calculated the Area Under the receiver operator characteristics Curve (AUC). The AUC informs on the accuracy of the intersectional strata information for discriminating those women who used antidepressants from those who did not. The AUC values extend from 0.5 to 1, where 0.5 represents absence of accuracy and 1 represents total accuracy. Both the VPC and the AUC in model 1 can be interpreted as measures of discriminatory accuracy,This model includes the fixed, main effects of all the intersectional dimensions  used to define the intersectional strata. In model 2, we quantified, (i) the association between the intersectional dimensions and use of antidepressants as expressed by OR and 95% CI. We also to calculate (ii) the proportional change in the variance (PCV). The PCV measures the overall proportion of strata variance of model 1 explained by the specific intersectional dimensions. Since model 2 contains all the variables used to construct the intersectional strata as main effects, it should explain all the strata variance . If this is not the case, the remaining between strata variance would be due to the existence of multiplicative interaction of effects between the intersectional dimensions defining the strata.The AUCs of the models 1 and 2 are expected to be the same because model 2 only decomposes the stratum-specific predicted probabilities obtained in model 1 into fixed and random-effect components and their sum equals the prediction obtained only by random effects in model 1.runmlwin command.We ran the models using MLwiN V.3.00 by calling it from within Stata V.14.1 using the The research was developed with a grassroot perspective in mind, whereby women’s experiences of use of HC inspired and informed the choice of research area and research questions. The anonymised data and scope of the study, including around 1 million women, prohibited direct patent involvement.The selection of the study population is shown in 10.1136/bmjopen-2021-049553.supp5Supplementary data10.1136/bmjopen-2021-049553.supp6Supplementary dataThe share of HC users was very similar in healthy women and those with previous mental health issues, 42.0% and 42.5%, respectively. Antidepressants were dispensed to 2.7% of HC users compared with 1.9% of non-users among healthy women during follow-up. For women with previous mental health issues, 41.2% of HC users and 39.8% of non-users dispensed an antidepressant prescription. The income levels were generally higher among women without mental health issues, and HC users were somewhat more affluent in both cohorts.Model 1 indicates that 8.45%  and 8.18%  of the total individual variance in the latent propensity of using antidepressant is at the intersectional strata level. These VPCs correspond with AUC values of 0.62 and 0.64, respectively. Both measures suggest the existence of a moderate intersectional effect. The PCV was high in both groups, but especially so in the group with previous mental health issues, meaning the intersectional dimensions or main effects explain more of the interstrata variance for these women. Model 2 shows that HC was associated with increased usage of antidepressants after adjustment for all other intersectional dimensions. This result was seen within both cohorts, but more strongly so in women without previous mental health issues (OR 1.62 compared with 1.19). Finally, the VPC in model 2 was very small  but did not vanish. This finding means that while the intersectional strata effect was mainly due the additive effect of variables defining the strata, a small component due to interaction of effects could also be detected.Women with previous mental health issues had a much higher usage of antidepressants than women without such issues, but the association with HC use nonetheless varied across the other intersectional dimensions. The highest use of antidepressants were observed in non-immigrant women, aged 24–30, with previous mental health issues, using HC and with low income (50.1%). The lowest usage were found in teenagers without previous mental health issues and no HC use, especially in the strata of immigrant girls from low (0.50%) and middle-income (0.60%) households.Overall, the propensity to use antidepressants was consistently higher in HC users compared with non-users in younger women between 12 and 17 years of age, both without previous mental health issues (0.7–2.4 percentage points), and with a mental health history (5.7–7.8 percentage points) with the magnitude being higher in the latter group. However, the 95% CIs were broad since the number of individuals was relatively small in these latter strata. The main hypothesis of our study was that the previously observed association between HC and antidepressant use, mainly seen in adolescent girls,selective discontinuation bias,healthy worker survivor effect, describing how bias is introduced through a continuous selection where those staying in the workforce are healthier than those who leave.selective discontinuation bias. This could explain why the observed association between HC and adverse mental health outcomes are stronger in adolescents. Most Swedish women do however continue their HC treatment with the same method.Independently of previous mental health issues, the propensity for using antidepressants was consistently higher for HC users than for non-users in teenagers aged 12–17, a result aligned with previous studies that has found a heterogeneous response with regard to both age and other factors.10.1136/bmjopen-2021-049553.supp7Supplementary dataAs expected, among adult women the overall propensity for using antidepressants was higher, as it is known that antidepressant use increases by age,10.1136/bmjopen-2021-049553.supp8Supplementary dataThe big difference in antidepressant consumption depending on HC use for lower income immigrant women could be interpreted as the intersectional contexts embodied by these women are more susceptible to the potential detrimental effect of HC on mood. The interrelating negative consequences of low income as a proxy for class or social position, gender and xenophobia may accumulate over the life course and lead to a higher vulnerability to exposures that predispose for antidepressant use later in life,Focusing on women whose lives are affected by several interlocking power dimensions such as low social position and xenophobia is fundamental to achieving reproductive justice.In this study, we have combined a classical epidemiological approach of exposure to HC and an intersectional MAHIDA to create a novel understanding of how intersecting power dynamics could create particular vulnerabilities to this specific exposure. Because of our study design, where women are followed for 1 year after a dispensed prescription of HC, it is more theoretically coherent to view use of HC as an exposure rather than a component of the intersectional strata. However, it is possible to within our approach view HC use as a sociocontextual factor that captures certain living conditions , which somewhat changes the interpretation of the results. This epistemological tension is not necessarily a limitation, but could enrich the dialogue in social epidemiology on whether it is possible to separate contextual factors from ‘pure’ exposure.The findings from this study must be interpreted in the context of its limitations. The SPDR has highly reliable data on dispensed prescriptions but cannot measure the actual use of dispensed medications. Whether the women was exposed to HC treatment during her entire follow-up is thus not possible to determine with our method, although previous Swedish data suggest continuation rates for any HC after 6 months are almost 90%.As in any observational study, ours only allows for measurements of associations and cannot determine causation. Furthermore, apparently strong average associations do not necessarily convey a high discriminatory accuracy (see elsewhere for a short review and discussion).It is important to recognise intersectional perspectives and interacting axes of oppression to tailor better public health interventions, as well as acknowledging the experiences of oppressed women to reach reproductive and social justice.


Our software is a useful tool for validating bioinformatics software and understanding the dynamics of the persistent HIV-1 reservoir.The complexities of viral evolution can be difficult to elucidate. Software simulating viral evolution provides powerful tools for exploring hypotheses of viral systems, especially in situations where thorough empirical data are difficult to obtain or parameters of interest are difficult to measure. Human immunodeficiency virus 1 (HIV-1) infection has no durable cure; this is primarily due to the virus’ ability to integrate into the genome of host cells, where it can remain in a transcriptionally latent state. An effective cure strategy must eliminate every copy of HIV-1 in this ‘persistent reservoir’ because proviruses can reactivate, even decades later, to resume an active infection. However, many features of the persistent reservoir remain unclear, including the temporal dynamics of HIV-1 integration frequency and the longevity of the resulting reservoir. Thus, sophisticated analyses are required to measure these features and determine their temporal dynamics. Here, we present software that is an extension of SANTA-SIM to include multiple compartments of viral populations. We used the resulting software to create a model of HIV-1 within host evolution that incorporates the persistent HIV-1 reservoir. This model is composed of two compartments, an active compartment and a latent compartment. With this model, we compared five different date estimation methods  to recover the integration dates of genomes in our model’s HIV-1 reservoir. We found that the Least Squares method performed the best with the highest concordance (0.80) between real and estimated dates and the lowest absolute error (all pairwise  Virus evolution is complex, from the population dynamics of epidemic spread to the complexities of the spread of a virus throughout the body of an individual host. Phylodynamic methods are essential for understanding these viral systems at all levels of epidemic hierarchies . However, such tools must be robustly tested before we have confidence in their efficacy to reconstruct viral evolution. A reliable way to test phylodynamic methods is to apply the tools to data sets where the results are confidently known. Since it is typically difficult to be certain of results derived from empirical data, it is imperative to validate phylodynamic methods on simulated data where inferences of parameter estimates can be compared with the truth.Human immunodeficiency virus 1 (HIV-1) infection can currently be managed via combination antiretroviral therapy (cART) by halting viral replication thereby lowering plasma viral load, prolonging progression to acquired immune deficiency syndrome (AIDS) and minimizing transmission risk . Howeverin silico simulation, but there are, to our knowledge, no tools currently available that are specifically designed to simulate HIV-1 genomes within host accommodating the HIV-1 reservoir. Current genome simulation software  except 1, that nucleotide position 9,167 is a guanine (G) instead of an adenine (A) to change the premature stop codon in HX-B2’s nef into tryptophan (W) and 2, the nucleotide at position 6,063 is a thymine (T) instead of a cytosine (C) to change the threonine (T) into a start codon. Genomes in the active compartment are subject to selection due to CD4 and HLA-I down-modulation in nef based on .The active compartment begins the simulation with 10 genomes that are copies of full-length  ancestral HIV-1 type B strain HX-B2   . This moOverall, the population size and genetic distance distributions were similar to the main 100 simulated data sets. Plots akin to the figures of the main article are shown in nef sequences sampled from an HIV-1-infected individual  from two time points while the individual was on suppressive cART. These sequences are available on GenBank with the following accession numbers: MG822918, MG822919, MG822923-MG822933, MG822935-MG822997, MG822999-MG823015, and MG823144-MG823170. More details on sample collection and sequencing can be found in Next, we compared our simulated data with data derived from an actual HIV-1 infected individual. We curated HIV-1 nef region and then we removed duplicated nef sequences keeping the earliest sampled sequence. We inferred ML phylogenies from the simulated nef sequences and the empirical nef sequences using RAxML  phylogenies. First, we clipped the simulated full-length genomes to the ng RAxML , creatinng RAxML , to maxi−3 nucleotide substitutions per site per year and the simulated data sets had a median evolutionary rate of 9.80 × 10−3 [IQR 8.02–10.9 × 10−3] nucleotide substitutions per site per year. Evolutionary rate was estimated via linear regression between the root to tip distances and collection dates of the active sequences.Phylogenies inferred from the simulated data displayed substantial variation in topology . In Fig.The phylogenies of the simulated data sets with varied parameters were within the scope of the phylogenies from the main 100 simulated data sets . Except nef sequences described in the previous section. Next, we assessed the accuracy of each method to investigate, which method is most appropriate for estimating the integration dates of latent HIV-1 proviruses.In HIV-1 persistence research, the timing of integration and duration of persistence of the HIV-1 reservoir is hotly contested. Three studies have attempted to resolve this debate , each usThe first method is Closest Sequence (CS), which entails assigning the date of a reservoir sequence based on the date of the phylogenetically closest active sequence. This method was one of three methods used by t tests: P < 0.01). The only method whose error had significant skewness was the CS method, which had a negative skewness (−1.31). Together with this method’s highly negative median error of −0.596 years indicates a tendency for this method to estimate older dates. The methods overall had negative skewness but were all greater than −1. The negative skewness of the methods may be a result of the distribution of the actual integration dates being skewed to younger dates and not an actual preference for estimating older dates.We applied the five date estimation methods to estimate the integration dates of the unique reservoir genomes to each of our 100 simulated data sets. The distributions of real and estimated integration dates are shown in There are many reasons why the LS method may outperform the other methods. Since the LS method treats time as a continuous variable, it is not restricted to the sampled time points for the estimated dates like in the CS and CD methods. The LR method assumes that the data follow a strict molecular clock, which is not held over longer periods of time during within-host HIV-1 infection , and theFinally, we investigated how varying the model parameters would affect the date estimation performance. The actual integration dates of the simulations all skew towards the start of therapy , consistnef codons: 123–143) to generate an alignment 63 nucleotide bases long for one data set. For comparison, nef is 618 bases long. We chose this region because it contained the most (four) amino positions under selection, according to our fitness model, than any other 63 nucleotide base long alignment. We used the full genome of the same data for a second data set. We removed duplicate sequences and inferred phylogenies from these alignments as previously described and performed the five date estimation methods. The results are shown in Supplementary Figures S26 and S27. The LR method had the highest concordance between estimated and actual dates (0.571) for the 63 base long alignment and the ML method had the highest concordance between estimated and actual dates (0.844) for the full genome alignment. The LS method did not perform much worse than the ML method in the full genome alignment with a concordance between estimated and actual dates of 0.834; however, the LS method performed poorly on the 63 base long alignment with a concordance between estimated and actual dates of 0.413. For context, the concordance between estimated and actual dates of the LS method on the original data set with a full nef alignment shown in To compare the effect of sequence length on the date estimation methods, we considered alignments with different numbers of bases. First, we clipped the simulated data shown in P-value <0.01. These results agree with the results from our in silico data, where the LS method performed the most accurately.Since simulations do not necessarily capture the entire complexity of the real world, we sought to compare the date estimation methods on our empirical data. We applied the five date estimation methods on the phylogeny derived from the patient data introduced in the “Comparison to empirical data” section to infer the integration dates of the proviral sequences. The date estimation methods for the empirical data were performed in the same manner as for the simulated data with the plasma-derived RNA sequences treated as coming from the active compartment and the PBMC-derived DNA sequences treated as coming from the latent compartment. Although we cannot be certain about the actual integration dates in the empirical data, we can compare the estimates from each method . In termHere, we present an extension of SANTA-SIM that enables the simulation of virus evolution in multiple compartments. Within this software we created simulated data sets of within host HIV-1 evolution including the HIV-1 reservoir. Our simulated data sets moderately resembled empirical data. Next, we utilized the simulated data sets to compare five date estimation methods to recover the estimated integration dates of reservoir genomes. Overall, we found that the LS method implementing LSD  yielded nef are far from the only evolutionary pressures faced by HIV-1. For example: cytokine, chemokine, SERINC3 and SERINC5 regulation  using R v3.6.2  that contains the query sequence and at least one sequence from the active compartment is selected. The query sequence is then assigned an integration date equal to the mode of the sampling dates of the sequences from the active compartment that are contained in the selected subtree. The CD method was implemented in a custom R script.10.3In the third method, a linear regression is inferred comparing the sampling dates versus the patristic distance from the root of the tree of the sequences from the active compartment. The integration dates of the sequences from the latent compartment are inferred from LR using their patristic distance from the root of the tree .10.4In the penultimate method, the dates of the internal nodes and sequences from the latent component of the tree are selected to minimize the variance between the branch lengths and the difference in time. LSD v0.3.3  was used10.5In the final method, the internal nodes and sequences from the latent compartment are assigned dates to maximize their likelihood. A modified version of node.dating available on GitHub (Jones and Poon 2016; veaa089_Supplementary_DataClick here for additional data file.


Reducing the consumption of meat constitutes an important part of the global shift towards more sustainable food systems. At the same time, meat is firmly established in the food culture of most human beings, and better understanding of individual behaviors is essential to facilitate a durable change in contemporary eating patterns. To determine the level and nature of attachment to meat among consumers, the Meat Attachment Questionnaire (MAQ) in relation to the phases of behaviour change in the meat consumption reduction process was utilised. Data collected through a survey carried out among Poles aged 25–40 years living in cities were analysed with the use of Spearman’s correlations and one-way ANOVA with Tukey’s post-hoc tests. The biggest share of the studied group of millennials (N = 317) never considered reducing their meat consumption (Phase 1–41%) and was described by the highest level of MAQ score in all its categories: hedonism, affinity, dependence, and entitlement. More than half of the respondents in Phase 2 participants (“planners”) declared a willingness to cut down meat consumption but had not yet put their intentions into practice. Respondents qualified in Phase 3 declared the highest willingness to reduce meat consumption and were significantly less attached to meat regarding all MAQ categories than respondents in Phase 1. The 9% of the study participants (Phase 4) had already limited the frequency of their meat consumption to “several times a week”, this however still remains insufficient compared to the ambitious goals of sustainable healthy diets. Results indicated that meat attachment categories, especially hedonism and dependence, were identified as predictors of willingness to reduce meat consumption. Research exploring the determinants of change and possibilities of effective communication about meat reduction on an individual level in different cultural settings are needed. The urgent need to change current dietary patterns has achieved a consensus around the world in the light of the global syndemic of obesity, undernutrition and climate change . ShiftinMany studies set meat in a central role in the development of humans, as a carnivorous species . InitialThe production of food of animal origin leads to significantly higher greenhouse gas (GHG) emissions per unit mass compared to the production of plant-based food raw materials. The results of the Global Livestock Environmental Assessment Model (GLEAM)  indicateGrowing concerns about unhealthy dietary patterns and food system sustainability are being mirrored in recommendations to reduce meat intake in developed countries. For example, the Danish dietary guidelines “good for health and climate” indicate that 350 g of meat per week is adequate in a plant-rich and varied diet and suggest especially limiting the consumption of beef and lamb . In PolaDietary guidelines in many countries (including Poland) advise eating more pulses and other foods that are sources of plant protein to achieve a food consumption levels that reflect a healthy sustainable diet ,37,38. TLimiting meat consumption is urgent and requires changes from all the stakeholders of the food system . The conIn our study we adapted two scales introduced in previous publications: phases of behavioral change based on Weibel et al.  and the A.What phase of change in terms of reducing meat consumption are the respondents in?B.Does willingness to reduce meat intake depend on socio-demographic variables?C.Is meat consumption frequency linked to the respondents’ self-assessment phase of change?D.Are behavioral phases linked to total MAQ scores and scores in each of the four categories of attachment?The attitudes of Polish 25–40 year-olds towards meat consumption was explored from the perspective of self-assessment variables and multidimensional background of attachment to meat. The study group was chosen due to the long-term nature of the effects of behaviour and dietary changes in younger adults. In a broader perspective, the choices of employees living in city influence the market and shape the food system as a whole, and at family level, as parents of small children, millennials, influence the formation of tastes and eating patterns of the next generation. Moreover, new consumer trends, including food and nutrition, spread from the inhabitants of large cities to the rural population, for which they constitute a certain model of future food consumption and behaviour . The folThe survey data were collected using the computer-assisted web interview (CAWI) method in March 2019. The questionnaire was distributed online among members of a consumer panel by a commercial market research company. Participants for the current study was recruited among individuals from another study published by our research team . Two criUltimately, 317 people took part in the study, including 161 women and 156 men, aged 25–40, working in cities in all 16 voivodships of Poland. The sample was of controlled quota type and was representative in terms of age and gender, based on demographic data from the national statistical office Statistics Poland (GUS). More than half (52%) of participants lived in towns and cities up to 100,000 residents, 27% in cities of 100–500,000 residents and 21% in the biggest Polish cities with more than 500,000 residents. White-collar workers consisted 70% of a sample, and blue-collar were 21%, 9% did not answer the question. The vast majority (82%) were employed under an employment contract, while 18% were employed on different kinds of contracts or self-employed. It was determined that 32% of the sample lived in three-person households, 24% in two-person households, 24% in four-person households, 12% lived alone, and 8% lived in households of five persons or more.The questionnaire covered many topics, not all of which were directly relevant to this paper. The survey’s parts used in this paper focused on meat-eating habits. The Meat Attachment Questionnaire was translated into Polish and verified by a bilingual speaker.To measure the frequency of meat consumption, a meat products index (MPI) was calculated as the arithmetic mean of the responses to two questions: the frequency of meat consumption and the frequency of meat product consumption  . Phase 2 participants (“planners”) declared that they had considered reducing their meat consumption but had not yet put their intentions into practice. Phase 3 respondents (“testers”) stated that “I make sure I consume less meat occasionally. In the future it is my firm intention to do this on a regular basis”. Those who indicated that they “take consuming little or no meat for granted” were grouped in Phase 4 (and labelled “reducers”). To verify the self-reported assessment of change, we asked whether respondents wanted to limit their meat consumption using a 5-point Likert scale from “1—definitely do not want”, to “5—definitely want”. Then, we correlated the results with the phases of change of meat consumption. In this part of the questionnaire, respondents indicated their level of agreement with 16 statements representing four categories of reasons for attachment to eating meat. Scores of each group of statements were averaged to create a category score, while all statements were averaged to create a total scale score. The hedonism category included four statements, where higher scores , D3: If I couldn’t eat meat, I would feel weak, D4: If I was forced to stop eating meat, I would feel sad and D5: I can’t picture myself not eating meat regularly. p < 0.001), and respectively between MPI and frequency of meat products consumption . Differences between respondents classified in the four phases and other variables were analysed using one-way ANOVA, with Tukey’s post-hoc tests. The internal reliability for the four MAQ categories was checked by the Cronbach’s alpha coefficient and was amounted to 0.882 (hedonism), 0.891 (affinity), 0.836 (entitlement), 0.851 (dependence).All statistical analyses were performed using the SPSS Statistics software package version 26 . The Spearman’s correlation was chosen to estimate the association between the created MPI and the frequency of meat consumption  of the sample was in Phase 1 (Ph1). Respondents in Phase 2 (Ph2) constituted 27% (n = 87) of the sample, 23% (n = 73) of individuals were in Phase 3 (Ph3) and 9% (n = 28) in Phase 4 (Ph4).The collected data showed that all four phases of behaviour change related to eating meat were represented in the study group. The biggest share, 41% (The meat products index and the declared phase of change in meat consumption were significantly correlated . Respondp < 0.001).More than 75% of regulars (Ph1) did not want to reduce the amount of meat they eat . On the p = 0.234) and there was also no relationship between phases and household size (p = 0.324). However, both characteristics determined the frequency of meat consumption (MPI). Males consumed meat more frequently than females (p = 0.046). The MPI for men was 3.77 and for women 3.54. The frequency of meat consumption increased with household size (p = 0.009). In one-person households, the MPI was 3.35, while in households of 4 or more people it was 3.89.No relationship was found between phases of meat consumption reduction and the gender of respondents  . Each stRegulars (Ph1) strongly agreed that meat is irreplaceable in their diet (mean 4.3) as well as that they cannot picture themselves not eating meat regularly (mean 4.1), which influenced the highest level of dependence category score compared to other phases. However, the reducers (Ph4) indicated the dependence opinions at the lowest level (mean 2.8) among all statements of the questionnaire. Consequently, testers and reducers (Ph3 and Ph4) were described by the lowest scores in the dependence category . The highest scores (mean 4.3) in all statements in the affinity category (with reversed scores) were obtained by regulars (Ph1). Surprisingly, reducers (Ph4) had the second highest mean score (3.8) in the affinity category. This paper points to a strong attachment to meat among people aged 25–40 years working and living in Polish cities, which is a predictor of resistance (low level of willingness) to cut back meat consumption. Respondents who intend to or have already reduced their meat consumption were driven by different levels of meat attachment factors compared to those with no willingness to change (and the differences were significant).In recent years excessive meat intake has become a more controversial issue due to social, environmental and ethical reasons. Despite extensive discussion and scientific evidence on the adverse consequences of eating meat, it is still widely consumed. People choosing to eat meat are often conflicted by the “meat paradox” ,54,55. TIt was also shown in our study that the dependence category divided consumers who declared regular meat consumption (Ph1) and those who evolved towards limiting meat intake (Ph3 and Ph4). The dependence category was related to the perception of meat being good for health. This link was shown also in other studies, where meat eaters were less likely to believe that meatless food choices are nutritionally adequate . Health One of the most relevant findings of our study is that there is a very different approach towards decreasing meat consumption among participants in each phase of behavioral change. Those individuals who declared eating little or no meat on daily basis, admitted eating meat at least several times a week. On the other hand, another study showed that consumers who considered themselves meat-eaters could restrict their meat consumption even to 1–2 times a week , which cThe respondents in our study were young people in their 20s and 30s. Making life-changing decisions that impact their lives for many years ahead is very common in this age group and may result in change of a lifecycle stage, which influences the attitudes towards food . With thTransitions in dietary behaviors requires behavioral facilitation and new value creation . ChanginThe present study has several limitations. Meat consumption was based on self-reported frequency of consumption so it might differ from an objective measure of actual meat intake due to individual concerns regarding social acceptance of this food group. Indicating the precise amount of meat consumed daily or weekly is crucial to define what meat reduction really means to consumers. The study sample was limited due to funding constraints to the respondents of a specific age and living in cities. We recommend further research in broader population groups, including inhabitants of rural areas in Poland and other countries. In order to explore the process of change and willingness to limit meat consumption it is important to understand the socio-cultural background of consumer behaviour. This study indicates that strong attachment to meat and especially its two categories—hedonism and dependence—can be predictors for a low level of willingness to decrease meat consumption among Polish millennials. However, the amount of meat consumed (even in the case of reduced amounts) is still high compared to the goals of a sustainable diet. The environmental and health domains of the concept share common aims, but social and cultural domains seem to remain in opposition to the necessary changes. There is a growing discussion in Europe about climate change, including the harmful effects of agriculture and, in particular, livestock production and consumption on planetary health. Knowledge about this topic and its impact on dietary behaviour is limited and requires further research in other population groups and countries where meat consumption is very high. The MAQ was found to be a useful tool for exploring meat attachment and when compiled with other measures, can provide profound research results that can further be used to develop sustainable food policy activities.Further research should include a more accurate description of the level and structure of meat consumption, both in terms of frequency and quantity, bridging together the MAQ tool with other measures in other population groups, including youth and rural inhabitants.