Interacting with others wearing a face mask has become a regular worldwide practice since the beginning of the COVID-19 pandemic. However, the impact of face masks on cognitive mechanisms supporting social interaction is still largely unexplored. In the present work, we focused on gaze cueing of attention, a phenomenon tapping the essential ability which allows individuals to orient their attentional resources in response to eye gaze signals coming from others. Participants from both a European  and an Asian  country were involved, namely two countries in which the daily use of face masks before COVID-19 pandemic was either extremely uncommon or frequently adopted, respectively. Both samples completed a task in which a peripheral target had to be discriminated while a task irrelevant averted gaze face, wearing a mask or not, acted as a central cueing stimulus. Overall, a reliable and comparable gaze cueing emerged in both experiments, independent of the mask condition. These findings suggest that gaze cueing of attention is preserved even when the person perceived is wearing a face mask. From the COVID-19 outbreak onwards, face mask has been introduced by governments worldwide as a protective gear to stop the spread of the virus e.g., . From a From a psychological perspective, faces have a profound impact on a variety of cognitive mechanisms supporting social interaction. For instance, from the unchangeable features of others’ face we can acquire information concerning their age, gender, and, more in general, identity . SimilarSocial attention has been widely studied by adopting the so-called gaze cueing task e.g., , which tTo sum up, in the present study, we conducted two experiments based on a gaze cueing task in which averted gaze faces, either wearing a face mask or not, were employed as cueing stimuli. The gaze cueing task was delivered to a sample of Italian individuals living in Italy (Experiment 1) and to a sample of Chinese individuals living in China (Experiment 2). Because ethnic membership can deeply shape gaze cueing e.g., , in bothData were collected from October 23, 2020, until November 7, 2020, during the second wave of the pandemic in Italy, which took place in the fall period. In particular, during the time window in which data were collected, the new daily cases of COVID-19 in Italy increased from 19.143 to 39.809, according to the COVID-19 data repository by N = 46 . Before the experiment, participants were asked to read an informed consent, which was provided by pressing a specific keyboard key. The project was conducted in line with the Declaration of Helsinki, and it was approved by the Ethics Committee for Psychological Research at the University of Padova.The sample was composed of Italian students at the University of Padova, Italy, who took part on a voluntary basis. Based on previous studies in which a whole face vs. only eyes were visible e.g., , we aimeFacial stimuli consisted of high resolution faces (300 px width × 450 px height), depicting two White males and two White females with neutral expression, extracted from the MR2 database . For eacIn the gaze cueing task . Before both the gaze cueing experiment and the questionnaire, a specific text screen appeared containing the instructions to successfully complete the requested task.https://www.jamovi.org/).Data of all participants who completed the experiment were analysed. Partial data provided by participants who aborted the task before its ending were not stored by the Pavlovia platform. Data analyses were performed through JAMOVI software  were rare and therefore not further analysed. Wrong responses  were analysed separately. Outliers, namely correct trials with a latency smaller than 150 ms or greater than 1500 ms , were discarded from RT analyses. Data were analysed both using a frequentist approach and a Bayesian framework, in order to establish the best model fitting the available data.F = 12.378, p < .001, 2pη = .216, due to smaller RTs on congruent  than on incongruent  trials, as well as the main effect of SOA, F = 125.698, p < .001, 2pη = .736, due to smaller RTs at the longer  than at the shorter  SOA, consistent with a foreperiod effect. No other results were significant , including the two theoretically relevant spatial congruency × mask, and spatial congruency × SOA × mask interactions  for being preferable over the null model, and strong evidence (10BF = 19.835) for being preferable over the first best model including the theoretically relevant interaction between spatial congruency and mask .RTs for correct trials were analysed through a repeated measures ANOVA, with spatial congruency (2: congruent vs. incongruent), SOA (2: 200 vs. 700 ms), and mask (2: mask vs. no mask) as within-participants factors. The main effect of spatial congruency was significant, F = 5.041, p = .030, 2pη = .101, with fewer errors at the longer  than at the shorter  SOA. No other results were significant . The Bayesian ANOVA, performed identically as that used for RTs analyses, confirmed that the model including only the main effect of SOA was the best one fitting the data. This model received anecdotal evidence (10BF = 2.228) for being preferable over the null model, and extreme evidence (10BF > 100) for being preferable over the first best model including the theoretically relevant interaction between spatial congruency and mask .Wrong responses were analysed through an identical ANOVA as that used for RT analyses. The main effect of SOA was significant, α = .20), they were kept separated; responses to the items concerning the subjective measures (Section 2 of the questionnaire) were intercorrelated (Cronbach's α = .74), and therefore they were averaged, thus obtaining a single score. Nevertheless, no significant correlations emerged for both the objective  and the subjective  measures. For completeness, correlations were also performed with an overall index of gaze cueing  and non-significant results emerged for both the objective  and the subjective  measures.The mean scores observed for each item of the questionnaire (see Appendix A) are reported, separately for each experiment, in Data were collected from November 11, 2020, until November 13, 2020. During the time window in which data were collected, the new daily cases of COVID-19 in China were extremely rare and changed from 15 to 18, according to the COVID-19 data repository .Mean age = 19.13 years, SD = .86, 14 males). The informed consent was provided by participants as in Experiment 1.We aimed to collect a similar number of participants as in Experiment 1. The sample was composed of 46 Asian students at the University of Guangzhou, China, who took part on a voluntary basis. Data collection was stopped at N = 46 (ps > .27).Everything was identical to Experiment 1, with only one exception: Faces belonged to Asian individuals. Importantly, since the MR2 database  providesData of all participants who completed the experiment were analysed. Partial data provided by participants who aborted the task before its ending were not stored by the Pavlovia platform. Data were analysed as in Experiment 1.Missing responses  were rare and therefore not further analysed. Wrong responses  were analysed separately. Outliers  were discarded from RT analyses.F = 55.335, p < .001, 2pη = .552, due to smaller RTs on congruent  than on incongruent  trials, as well as the main effect of SOA, F = 83.247, p < .001, 2pη = .649, due to smaller RTs at the longer  than at the shorter  SOA. The spatial congruency × SOA interaction was also significant, F = 6.047, p = .018, 2pη = .118, and it was further explored, for each SOA, through two tailed paired t-tests between congruent and incongruent trials. These indicated that the gaze cueing effect emerged both at the 200-ms SOA, t(45) = 3.342, p = .002, d = .493, and at the 700-ms SOA, t(45) = 6.610, p < .001, d = .975, but it was greater in the latter case (14 vs. 30 ms). No other results were significant , including the two theoretically relevant spatial congruency × mask, and spatial congruency × SOA × mask interactions  for being preferable over the null model, and very strong evidence (10BF = 40.087) for being preferable over the first best model including the theoretically relevant interaction between spatial congruency and mask .RTs of correct trials were analysed through a repeated measures ANOVA, with congruency (2: congruent vs. incongruent), SOA (2: 200 vs. 700 ms), and mask (2: mask vs. no mask) as within-participants factors. The main effect of spatial congruency was significant, F = 16.955, p < .001, 2pη = .274, with fewer errors at the longer  than at the shorter  SOA. No other results were significant . The Bayesian ANOVA confirmed that the model including only the main effect of SOA was the best one fitting the data. This model received extreme evidence (10BF > 100) for being preferable over the null model, and extreme evidence (10BF > 100) for being preferable over the first best model including the theoretically relevant interaction between spatial congruency and mask .Wrong responses were analysed through an identical ANOVA as that used for RTs analyses. The main effect of SOA was significant, α = .44) and the average index for subjective measures (Cronbach's α = .76) did not show any significant result . This holds true also for the overall index of gaze cueing .Mean scores collected through the questionnaire are reported in Fs > 61.508, ps < .001), as well as the spatial congruency × SOA interaction, F = 7.758, p = .007, 2pη = .079, reflecting that the gaze cueing effect, albeit significant at both SOAs , was greater at the longer one. Moreover, the main effect of experiment was not significant, F = .461, p = .499, 2pη = .005, but the experiment × congruency interaction was significant, F = 9.301, p = .003, 2pη = .094, reflecting that the gaze cueing effect, albeit significant in both samples , was greater among Asian than European participants . More importantly, experiment was not involved in any interaction including both spatial congruency and mask , thus indicating a similar gaze cueing effect in response to faces irrespective of whether wearing a mask or not, in both samples. The Bayesian ANOVA, including the same factors as those reported for the frequentist approach, indicated that the best model fitting the data included the main effects of spatial congruency, SOA, and experiment, as well as the spatial congruency × SOA and spatial congruency × experiment interactions. This model received extreme evidence (10BF > 100) for being preferable over the null model, and extreme evidence (10BF > 100) for being preferable over the first best model including the theoretically relevant interaction between spatial congruency and mask .In order to explore whether the gaze cueing effect was different between the two samples, we conducted a mixed design ANOVA, with the same within-participants factors used in previous analyses , and with experiment (2: Experiment 1 vs. Experiment 2) as an additional between-participants factor. Both the main effects of spatial congruency and SOA were significant (d = 0.058) suggests that regardless of the statistical significance, the difference, if any, would reflect a practically-irrelevant effect.As a final note, the effect size of the difference between the magnitude of gaze cueing in the mask and the no mask condition , From a purely methodological point of view, social attention abilities have been widely explored by adopting gaze cueing tasks based on manual responses, such as the one employed here. However, in every day social interactions, we tend to explore the surrounding environment around us through eye movements, which can be considered as a more direct and ecological index of visuo-spatial orienting of attention e.g., , and canOne of the main concerns regarding face mask use is related to the impact that such protective gear can have on interpersonal communication e.g., . Indeed,


Laboratory testing showed elevated plasma lactate and 3‐methyl‐glutaconic and tricarboxylic aciduria. Supplementation with high dose of coenzyme Q10 was not effective to control disease progression and the patient died at the age of 3 years old because of a progressive multisystem disorder. Cutaneous involvement in mitochondrial disease is heterogenous, including proliferative, inflammatory, and dystrophic changes among others. The coexistence in our case of phalangeal erythema, livedo reticularis, and pulmonary hypertension suggests microvascular dysfunction as a possible underlying mechanism. This is the first reported patient with PDSS1 mutations presenting with 3‐methyl‐glutaconic aciduria and distal phalangeal erythema, expanding the phenotype of primary coenzyme Q10 deficiency.We report a detailed clinical examination in a patient with primary coenzyme Q At this time, oral ubidecarenone was started at 15 mg/kg/day. Three months later, when biallelic PDSS1 mutations were identified  and NM_014317.3: c.1183C > T (p.Arg395*)), ubidecarenone was increased to 30 mg/kg/day.At the age of 5 months, he presented failure to thrive, hypotonia, feeding difficulties, bilateral hearing impairment, pulmonary hypertension, livedo reticularis, and distal phalangeal erythema Figure . LaboratHowever, he developed a progressive multisystem disorder with severe developmental delay, pyramidal dysfunction, seizures, tremor, optic atrophy and nystagmus, severe visual and hearing impairment, white matter cystic change and chronic heart failure. He died at the age of 3 years old.10 deficiencies,PDSS1,COQ4,CoQ8APDSS1 variants and primary coenzyme Q10 deficiency.Since the age of 5 months, distal phalangeal erythema was an accompanying sign. Cutaneous involvement in mitochondrial disease is heterogenous, including proliferative, inflammatory, and dystrophic changes among others.Marcello Bellusci, Maria Teresa García‐Silva, Ana Martínez de Aragón, and Miguel Angel Martín declare that they have no conflict of interest.Marcello Bellusci and Maria Teresa García‐Silva are the clinician who attended the reported patient, planned, wrote and revisited critically the article before the submission. Ana Martínez de Aragón is the pediatric radiologist that realized the reported brain MRI studies, wrote the figure legend and revisited critically the article before the submission. Miguel Angel Martín performed and analyzed laboratory tests, revisited critically the article before the submission.All procedures followed were in accordance with the ethical standards of the responsible committee on human experimentation  and with the Helsinki Declaration of 1975, as revised in 2000 (5). Informed consent was obtained from all patients for being included in the study, including for the use of pictures.No ethics approval is required for case report in our center.


Educational attainment is moderately heritable and inversely associated with the risk of rheumatoid arthritis. However, the causality from educational attainment on rheumatoid arthritis remained unknown. Here, we aimed to determine whether educational attainment is causally associated with rheumatoid arthritis (RA) by using Mendelian randomization (MR) approach.N = 1,131,881) of European ancestry. MR analyses were mainly performed using the inverse-variance weighted (IVW) method. Sensitivity analyses were further performed to test the robustness of the association using the weighted median method, MR-Egger, Cochran Q test, “leave-one-out” analysis and MR-PRESSO test.Summary statistics data for RA were obtained from an available, published meta-analysis of genome-wide association studies (GWAS) that included 14,361 RA cases and 43,923 controls of European ancestry. The instrumental variables for educational attainment were obtained from a GWAS meta-analysis that included over 1 million individuals (p = 1.78 × 10− 14). The weighted median method and MR Egger regression analysis yielded consistent results. The effect estimate remained robust after the outlier variants and SNPs (associated with the confounding factors) were excluded. “Leave-one-out” analysis confirmed the stability of our results. Additionally, the results suggested the absence of the horizontal pleiotropy.A total of 387 SNPs were employed as instrumental variables in our MR analysis. Genetically predicted higher educational attainment was associated with a significantly lower risk of RA using the IVW method (odds ratio [OR] = 0.42, 95% confidence interval [CI]: 0.34–0.52; The MR analysis supported a potential inverse causative relationship between educational attainment and the risk of RA.The online version contains supplementary material available at 10.1186/s41927-021-00216-0. Rheumatoid arthritis (RA) is a systemic autoimmune disease with a lifetime prevalence of 0.5 ~ 1% worldwide . And it’Although the exact cause of RA remains unclear, both environmental and genetic factors contribute to development of the disease. Previous observational epidemiological studies have shown an inverse association between educational attainment and the risk of RA. However, this association may be blurred by the methodological limitations of traditional observational studies, including residual confounding, reverse causation, and measurement error .Educational attainment is a well-established socioeconomic and heritable determinant of health , which iMendelian randomization (MR) has been widely used to evaluate causality by exploiting genetic variants with SNPs as instrumental variables to predict the effect of an exposure on a particular outcome , which oThis study applied MR as a method to determine whether educational attainment is causally associated with RA, using summary data of SNP-exposure  and SNP-outcome (rheumatoid arthritis) based on genome-wide association studies (GWAS). An overview of the study design is shown in Fig. Assumption 1, genetic variants should be strongly associated with the exposure; Assumption 2, genetic variants extracted for exposure should be independent of any confounder which is associated with both exposure and outcome; and Assumption 3, the genetic variants affect the outcome only through the exposure. *Sensitivity analyses: Weight median method, MR-Egger regression, MR-PRESSO, Leave-one-out test.Publicly available summary statistic estimates for the associations between genetic variants and risk of RA were obtained from a GWAS meta-analysis, including 58,284 individuals from 18 studies of European ancestry  . All RA N = 1,131,881). All cohort-level analyses were restricted to European-ancestry individuals who passed the cohort’s quality control and whose educational years were measured at an age of at least 30. The phenotype was constructed by mapping each major educational qualification that can be identified from the cohort’s survey measure to an International Standard Classification of Education (ISCED) category and imputing a years-of-education equivalent for each ISCED category. In this study, the educational attainment was evaluated by years of schooling completed  = 16.8 ± 4.2 years).The genetic instrumental variables associated with educational attainment were obtained from a GWAS meta-analysis  comprisip < 5 × 10− 8). After clumping correlated SNPs (linkage disequilibrium [LD] r2 ≥ 0.001), 393 SNPs remained and were used as instrumental variables. We then extracted educational attainment–associated SNPs from the outcome data . For SNPs absent in the outcome data, we identified proxy SNPs at a cutoff of LD of r2 > 0.8 from the SNiPA website (https://snipa.helmholtz-muenchen.de/snipa3/index.php). SNPs missing in the outcome data without appropriate proxy SNPs available were then excluded. We then calculated the F statistic for each of the SNPs using the following formula: R2 × (N − 2)/(1 − R2) . Here, R2 indicates the proportion of variance in educational attainment explained by a given SNP and N indicates sample size. More specifically, R2 was calculated with the following formula: R2 = [2 × Beta2 × (1 − EAF) × EAF]/[2 × Beta2 × (1 − EAF) × EAF + 2 × SE2 × N × (1 − EAF) × EAF]. Here, Beta indicates the genetic effect of SNP on educational attainment, EAF is effect allele frequency, SE is standard error and N is sample size. F statistic is recommended to be over 10 to avoid employing week genetic instruments [A number of quality control steps were taken in our analysis to select eligible instrumental SNPs that were strongly associated with educational attainment. First, we identified 1271 lead SNPs at the genome-wide significance threshold (truments .https://www.r-project.org/); p < 0.05 was the threshold for a significant difference. All estimates were reported with two-tailed p-values. In the main analysis, we utilized the inverse-variance weighted (IVW) method to investigate the causality between educational attainment and RA [Statistical analyses were performed using the two-sample MR package (version 0.5.5) in R software version 4.0.2  for whether any SNP in this study was associated with these confounders at the genome-wide significance of p < 5 × 10− 8. Then we found 18 SNPs associated with confounders . These risk factors might be potential confounding factors of this MR study. And confounders considered in this study should be associated with both RA and educational attainment. Therefore, we further conducted a comprehensive search in the GWAS Catalog  and 393 SNPs remained after clumping (r2 < 0.001). However, seven SNPs were not available in the summary statistics data for RA. Of these, one proxy variant  was identified for the missing SNP (rs72486027). Six SNPs  without appropriate proxy SNPs available were excluded. And after the harmonizing process, one SNP (rs77719387) was removed for incompatible alleles and 13 SNPs were excluded for being palindromic with intermediate allele frequencies . Therefore, 373 SNPs were chosen as instrumental variables for educational attainment in the present study . The SNP rs2256965 was the only outlier variant detected by the MR-PRESSO test.A total of 1271 lead SNPs were identified at the genome-wide significance threshold (p = 1.78 × 10− 14), which meant that RA risk decreased by 58% per SD (approximately 4.2 years) increased in the years of schooling. The associations were consistent in the sensitivity analysis using the weighted median method . The effect was only slightly attenuated using the MR-Egger method . Considering that the weighted median estimator had the advantage of retaining greater precision of the estimates compared with the MR-Egger analysis [p = 4.89 × 10− 16).As were shown in Fig. analysis  and MR-Ep = 1.82 × 10− 12. However, after removing the outlying SNP, heterogeneity was remarkably decreased: Q value (df) = 425.7(371), p = 0.03. Additionally, we found no indication of unbalanced pleiotropy . The results of the leave-one-out analysis  demonstrated no potentially influential SNPs driving the causal link between educational attainment and RA in the replicated analyses.Heterogeneity tests suggested an apparent sign of heterogeneity: Q value (df) = 594.21(372), p = 1.05 × 10− 15, using the IVW method).We then scanned the SNPs for their potential secondary phenotypes using the GWAS catalog. As was shown in Additional file, Table 4, a total of 18 SNPs associated with educational attainment were found to be associated with other traits affecting RA. After excluding SNPs associated with potential confounders, results from the statistical analysis remained essentially consistent  as the exposure and a meta-analysis of GWAS of RA (n = 5539) and European controls  as the outcome. Here, we expanded the exposure sample size to over 1 million individuals  and chose the latest meta-analysis of GWAS, which included 58,284 individuals of European ancestry . Meanwhile, the number of SNPs chosen as instrumental variables increased dramatically (from 49 to 373). Furthermore, individuals with RA who were seropositive and seronegative for ACPA or RF were enrolled in this MR analysis. Recently, a comparable analysis has been carried out by Yuan et al. used the same GWAS as sources for educational attainment  and rheumatic arthritis  [2 of the SNPs in our study was set to a more conservative threshold to obtain a smaller set of SNPs (LD r2 < 0.001 with 387 SNPs remained vs. LD r2 < 0.01 with 663 SNPs used in Yuan’s study) to ensure the independence of SNPs at the cost of decreasing statistical power. Secondly, the RA phenotype used in Yuan’s study came from mixed populations (European and Asian ancestry). However, it should be noticed that population heterogeneity may lead to bias to the MR results. As such, all the association analyses performed in our study were restricted to European-descent individuals, making the MR estimates more reliable. As such, our MR study is based on merely European descents and a set of more conservative genetic instruments, to appraise the causal relationship between EA and the risk of RA. Thus, the causative association fully explored in patients between educational attainment and risk of RA was more convincing.A previous two-sample MR study conducted by Bae and Lee  suggestes N = 1,1,881 and In total, the identified exposure SNPs accounted for approximately 11% of the variance in educational attainment. The effect size of the independent SNPs corresponding to an educational increase was obtained as follows: the median effect size corresponded to 1.7 weeks of schooling per allele (95% CI: 1.1–2.6 weeks). Furthermore, the genes related to these SNPs are involved in almost all aspects of neuron-to-neuron communication . The draThe MR approach, as an approximation to a randomized controlled trial in nature, offers one of the most compelling methods to detect causation. The IVW method and weighted median method suggested an inverse causal association between educational attainment and RA, whereas the MR-Egger method showed no proof of a causative association between educational attainment and RA. However, the MR-Egger test provided a reference for the direction of causal association. The weighted median method, which is not influenced by outlying genetic variants, improved the power of causal effect detection and effectively decreased type I error . Therefop = 1.82 × 10− 12). However, heterogeneity was decreased after removing the outlying SNP detected by MR PRESSO test  = 425.7(371), p = 0.03). Additionally, there was no indication of unbalanced pleiotropy . Therefore, we deemed that the conclusion would not be biased significantly by the heterogeneity of the analysis because several robust methods were performed, which could provide reliable inferences and statistical support.The results of our MR analysis might be biased by pleiotropy. Heterogeneity tests suggested an apparent sign of heterogeneity  = 594.21(372), The potential mechanisms that educational attainment ultimately reduces the risk of RA may be complicated. In general, higher educational attainment is associated with greater wealth and status, as well as healthier lifestyle and relatively higher quality of life. This cascade of benefits from higher educational attainment may eventually contribute to RA prevention. In addition, the effect of educational attainment on RA may also be mediated by obesity. In an MR study, Böckerman et al. found that the higher years of schooling was associated with lower BMI . AnotherThis study has several limitations. The summary GWAS data were restricted to individuals of European descent, and, because ethnicity may affect causality, our results may not be fully representative of the non-European populations. Another limiting factor was that this applied analysis could not be stratified by gender and age due to the meta-GWAS was performed without adjustment for gender or age to maximize statistical power, thus, we could not assess gender or age discrepancies and potential nonlinear associations.In conclusion, our aim in this study was to assess the causal effect of educational attainment and risk of RA by using two-sample MR analysis with pretty large sample sizes. However, further confirmatory methods should be conducted to verify our findings of a potential causal association between increased educational attainment and lower risk of RA. These results advocate the current clinical practice for RA surveillance in those with lower educational attainment.Additional file 1: Table S1. Detailed information of SNPs Harmonizing. Table 2. Detailed information of LD-independent SNPs chosen as instrumental variables for educational attainment (exp) and rheumatoid arthritis (out). Table 3. Detailed information of the “leave-one-out” analysis corresponding to the IVW analysis. Table 4. The SNPs and their corresponding phenotypes considered as confounders.


The COVID‐19 pandemic has been associated with excess mortality and reduced emergency department attendance. However, the effect of varying wave periods of COVID‐19 on in‐hospital mortality and length of stay (LOS) for non‐COVID disease for non‐COVID diseases remains unexplored.We examined a territory‐wide observational cohort of 563,680 emergency admissions between January 1 and November 30, 2020, and 709,583 emergency admissions during the same 2019 period in Hong Kong, China. Differences in 28‐day in‐hospital mortality risk and LOS due to COVID‐19 were evaluated.p‐interaction < 0.001). The overall average LOS in the pandemic year was significantly shorter than that in 2019 . Patients with mental disorders and cerebrovascular disease in 2020 had a 3.91‐day and 2.78‐day shorter LOS than those in 2019, respectively.The cumulative incidence of 28‐day in‐hospital mortality increased overall from 2.9% in 2019 to 3.6% in 2020 . The aHR was higher among patients with lower respiratory tract infection (aHR: 1.30 95% CI 1.26 to 1.34), airway disease (aHR: 1.35 95% CI 1.22 to 1.49), and mental disorders (aHR: 1.26 95% CI 1.15 to 1.37). Mortality risk in the first‐ and third‐wave periods was significantly greater than that in the inter‐wave period (Increased risk of in‐hospital deaths was observed overall and by all major subgroups of disease during the pandemic period. Together with significantly reduced LOS for patients with mental disorders and cerebrovascular disease, this study shows the spillover effect of the COVID‐19 pandemic. They also play a vital role in facilitating efficient use of resources by reducing pressure on in‐patient beds while preventing treatment delay and reducing mortality with limited resources. Before the pandemic, studies have been reported that ED crowding increased length of stay (LOS) and in‐hospital mortality.Given the necessity to maximize efficient use of resources and quality of patient care in hospital, information on patterns of ED admission, LOS, and mortality,Previous research has established that a decreased hospital admission and a higher mortality rate for non‐COVID medical conditions during the pandemic compared with the pre‐pandemic period, especially for patients admitted with respiratory diseases, cancer, pneumonia, and sepsis.22.1This retrospective study was based on entries in the territory‐wide electronic health care database, the Clinical Data Analysis and Reporting System (CDARS) of the Hospital Authority, Hong Kong SAR, China. The Hong Kong Hospital Authority managed all public hospitals and mobile clinics in Hong Kong. These public hospitals with EDs supplying a population exceeded 2 million in 2019,2.2All patients admitted as emergencies in Hong Kong between January 1, 2019, and November 30, 2019, and between January 1, 2020, and November 30, 2020, were included for analysis. We excluded hospital episodes with missing data for LOS, which represented a limited proportion of total ED admissions (0.4%). Patients who visited EDs (EDs visit) and were then admitted to hospitals were regarded as ED admissions and were included in our study as identified by records of destination of discharged at EDs. Patients who were hospitalized in public hospitals, whether transferred or directly admitted to other departments of that hospital from EDs, were included.Our study included a list of patients' covariates, comprising age, gender, race (Chinese and non‐Chinese), living regions , living in a residential care home for the elderly, and recipients of comprehensive social security assistance (CSSA). Severity of cases was defined by the mode of arrival, time of attendance , triage category, total time spent in ED (≤4 h or not), and disease diagnosis. The International Statistical Classification of Disease and Related Health Problems, Ninth Revision (ICD‐9‐CM) diagnosis codes were used to identify primary diagnosis of the following diseases: lower respiratory tract infection (LRTI), airway disease, hypertension, coronary heart disease (CHD), cerebrovascular disease and myocardial infarction (MI), diabetes, mental disorders, sepsis, chronic kidney diseases, cancer, and COVID‐19‐related conditions. The specific diagnosis for admitted patients was recorded during hospitalization other than diagnosis from ED. A social deprivation index (SDI), an indicator of socio‐economic status, was measured for each district by averaging six related variables.The first confirmed COVID‐19 case in Hong Kong was announced on January 23, 2020.2.3The primary outcome was the time from emergency admission to in‐hospital death within the 28‐day follow‐up period. The original definition of 28‐day in‐hospital mortality included all deaths that occurred during hospitalization within 28 days but excluding deaths that occurred after discharge from hospital. The International Statistical Classification of Diseases and Related Health Problems, Tenth Revision (ICD‐10) diagnosis codes were used to identify the primary cause of death. Diagnosis codes are listed in Table 2.4Descriptive statistics of patients' covariates including demographic and clinical characteristics, and the LOS of each hospital episode in 2019 and 2020 were displayed separately. Balance of patients' covariates between the 2 years was evaluated using the absolute standardized mean difference (SMD). An SMD <0.1 implied a covariate balance between the years.The incidence rate was a more precise estimate of the true risk of death in the dynamic populations.In order to examine the heterogeneous effects of COVID‐19 on in‐hospital mortality across the years and waves, the primary end point of time from emergency admission to in‐hospital death before and after pandemic or during different pandemic waves in 2020 was assessed by a Kaplan–Meier plot and compared with a log‐rank test. Cox proportional hazard models were performed to assess the impact of COVID‐19 on incidence of in‐hospital death within 28 days. Each hospital episode was observed from the date of emergency admission to one of the following events: discharge or 28 days as right censoring. Hazard ratio (HR) with 95% confidence interval (CI) for 2020 versus 2019 adjusted by individual and hospital characteristics was reported for the overall study population and the different subgroups. Proportional hazard assumptions were tested by using a log–log plot for categorical variables. There was no evidence to reject the proportional hazards assumption.For the secondary outcome measurements, the mean differences in LOS between the 2 years were generated by linear regression after adjusting for individual and hospital characteristics. Patients who died on the day of hospital admission would have a follow‐up time and an LOS of 0.5 days for consistency with another study,p value < 0.05 was regarded as statistically significant. All data management and statistical analyses were performed by using STATA version 16.0 .A 3n = 709,583), there was a 20.6% reduction in hospital admissions in 2020 ; 2.9%  and 3.6%  of these patients eventually died within 28 days of emergency admission during hospitalization in 2019 and 2020, respectively. The mean LOS decreased slightly from 5.49 days in 2019 to 5.17 days in 2020 (p < 0.001).A total of 1,273,263 emergency admissions were included during the 2‐year study period. Compared with 2019  . A significantly greater effect of the COVID‐19 pandemic on in‐hospital mortality was observed in the first wave and the third wave (p‐interaction < 0.001) compared with the interwave period. The increase in 28‐day in‐hospital mortality in 2020 was significant, especially among patients admitted with airway disease (aHR: 1.35 95% CI 1.22 to 1.49), LRTI (aHR: 1.30 95% CI 1.26 to 1.34) and mental disorder (aHR: 1.26 95% CI 1.15 to 1.37).Figure Figure 44.1This population‐based cohort study reported the time course of 28‐day in‐hospital mortality and LOS among patients admitted as emergencies before and during the COVID‐19 pandemic. 20,502 (3.6%) and 20,519 (2.9%) people died 28 days within the hospital in 2020 and 2019, respectively, while a sharp decrease (20.6%) in emergency admissions during pandemic year may have contributed to a higher in‐hospital mortality rate. In 2020, significantly higher 28‐day in‐hospital mortality rates after adjusting were observed overall and across a set of major diagnosis groups such as LRTI, airway disease, and mental disorders during the pandemic period compared with 2019. Although a minor difference of LOS in overall before and after COVID‐19 pandemic years, LOS greatly decreased for some medical conditions, for example, mental disorders, cerebrovascular disease, sepsis, and cancer.4.2The nature of the impact of COVID‐19 on in‐hospital mortality reported from previous studies has not been consistent, probably because of differences in the data source, local government intervention, study design, and statistical methodology. Plus, the degree of pressure that hospitals have come throughout the pandemic might be another major factor. Some studies found an increase in in‐hospital mortality in 2020,On one hand, finding a balance between the urgency of medical care and the perceived risk of infection with COVID‐19 results in decreased emergency admissions. Unnecessary hospitalization may have been prevented to some extent because physicians have to rethink which treatment is truly necessary at circumstance of limited medical resource during pandemic.The trend towards a significant increase in in‐hospital mortality in the third wave is the more worrying. Unlike the similar increase patterns for in‐hospital mortality and confirmed COVID‐19 cases for the first wave, the in‐hospital mortality grows gradually after the beginning of the third wave in all major diagnosis groups Figure . A U.S. Notably, our findings show an increased risk of death in patients with non‐COVID‐related respiratory disease and mental disorders in 2020. Previous studies have reported that patients with cardiovascular conditions, cancer, and respiratory disease were considered at higher risk of death during the pandemic.Our study found statistically significant decrease in LOS during pandemic year, and the slight difference of mean LOS before and after pandemic suggests this result might not be practically significant. However, the large reduction in LOS among patients who die in hospital is most marked in those who died within 2 days after admission Figure . These aThis study addressed two important research gaps. First, by using the Hong Kong Hospital Authority's comprehensive and reliable database, we could examine the association of in‐hospital mortality with different clinical groups. In‐hospital mortality rates following emergency admission have varied across countries and health care jurisdictions.4.3However, the study has limitations. First, this is an observational study that inherited the nature of residual confounding and might not deduce the causal effects of COVID‐19 pandemic in different waves on in‐hospital mortality and LOS. However, due to small differences in measured covariates between the 2 years, increased in‐hospital mortality and decreased LOS in year 2020 were likely attributable to COVID‐19. Second, laboratory parameters, comorbidities, drugs administered, operations, and procedures performed on our cohort of emergency admissions and the utilization of the negative pressure isolation rooms were not available for this analysis. We are not able to evaluate the impact of COVID‐19 on the out‐of‐hospital death and the degree of pressure on specific specialties in hospital. However, for the variables that were available, we had over 97% complete data. Third, our data only included patients admitted through EDs and represent a specific component of hospital admissions. In general, nonemergency admitted patients have mild and less complex medical conditions5In 2020, patients admitted as emergencies experienced a higher 28‐day in‐hospital mortality rate and shorter LOS in Hong Kong. The trend towards a significant increase in in‐hospital mortality in the third wave shows the spillover effect of COVID‐19. Increased risk of in‐hospital deaths was observed for respiratory disease and mental disorders during the pandemic period. Together with significantly reduced LOS for patients with mental disorders and cerebrovascular disease, the COVID‐19 pandemic may result in long‐term increases in mental health problems. The results of this study provide a useful reference when preparing health precaution campaigns and for forecasting healthcare resources needed during a pandemic.Xi Xiong: Data curation; formal analysis; investigation; methodology; validation; visualization. Abraham K.C. Wai: Data curation; investigation; project administration; supervision; validation. Eric H.M. Tang: Formal analysis; methodology; visualization. Owen C.K. Chu: Data curation; formal analysis. Carlos K.H. Wong: Conceptualization; investigation; methodology; project administration; software; supervision; validation; visualization. Timothy H. Rainer: Conceptualization; investigation; project administration; supervision; validation; visualization.This study was approved and a waiver of the need for patient consent allowed by the institutional review board of the University of Hong Kong/Hospital Authority West Cluster (UW 20‐112).https://publons.com/publon/10.1111/irv.12919.The peer review history for this article is available at Table S1. Primary disease diagnosis and causes of death defined by International Classification of Diseases, Ninth Revision, Clinical Modification (ICD‐9‐CM) diagnosis codes and International Classification of Diseases and Related Health Problems, Tenth Revision (ICD‐10) diagnosis codes.Table S2. Number of hospital admission at risk and the percent decreased in 2020 per week.Table S3. Demographics, clinical characteristics of emergency admissions by waves among emergency admissions on 1 January – 30 November 2019 and 1 January – 30 November 2020Table S4. 28‐day in‐hospital mortality overall and by causes of death by waves among emergency admissions on 1 January – 30 November 2019 and 1 January – 30 November 2020Figure S1. Kaplan–Meier survival curve for cumulative in‐hospital mortality within 28 days hospitalization by years (A) and waves (B)Figure S2. The distributions of LOS for patients with different discharge status by waves and years..Figure S3. Log hazard ratio  of 28‐day in‐hospital mortality among emergency admissions in 2020 compared with 2019Click here for additional data file.


Daily and event‐driven PrEP are both efficacious in reducing the risk for HIV infection. However, the practice of event‐driven PrEP (edPrEP) is less well studied, in particular when provided as an alternative to daily PrEP. We studied regimen preferences and switches, and sexually transmitted infection (STI) incidence.We analysed pooled data from two prospective cohort studies among MSM: Be‐PrEP‐ared, Belgium and AMPrEP, the Netherlands. In both projects, participants could choose between daily and edPrEP at three‐monthly study visits, when they were also screened for sexually transmitted infections including hepatitis C (HCV). We assessed the proportion choosing each regimen, and the determinants of choosing edPrEP at baseline. Additionally, we compared the incidence rates (IRs) of HCV, syphilis and chlamydia or gonorrhoea between regimens using Poisson regression. The study period was from 3 August 2015 until 24 September 2018.We included 571 MSM, of whom 148 (25.9%) chose edPrEP at baseline. 31.7% of participants switched regimen at least once. After 28 months, 23.5% used edPrEP. Older participants (adjusted odds ratio (aOR) = 1.38 per 10 years, 95% confidence interval (CI) = 1.15 to 1.64) and those unemployed  were more likely to initially choose edPrEP. IR of HCV and syphilis did not differ between regimens, but the IR of chlamydia/gonorrhoea was higher among daily users .A quarter of participants chose edPrEP at baseline and at 28 months this proportion was similar. Although the IR of HCV and syphilis were similar in the two regimens, the lower incidence of chlamydia and gonorrhoea among edPrEP users may suggest that less frequent STI testing of this group could be considered. A22.1We used pseudonymized pooled data from the Be‐PrEP‐ared and AMPrEP studies. Pooling data enabled us to study a relatively large number of edPrEP users and thus increase statistical power. Full procedures of both studies have been published previously , 20, 22.2.2At baseline, participants completed self‐administered computer‐assisted questionnaires with questions on socio‐demographic characteristics (e.g. education) and sexual behaviour.Details of STI screening and laboratory procedures have been described previously , 23, 24.In the Be‐PrEP‐ared study, hepatitis C (HCV) antibodies (anti‐HCV) were initially tested at the screening visit and after 18 months of follow‐up. HCV testing frequency was increased to six‐monthly starting May 2017 , 25. Add2.3χ2 or Fisher’s exact tests for categorical variables. Determinants for choosing edPrEP at baseline were assessed using logistic regression with a random‐intercept to account for between‐city variability. In multivariable analysis we included determinants associated in univariable analysis at p < 0.20 . Backwards selection was performed to obtain a parsimonious multivariable model.Baseline demographics and PrEP choice were compared between Be‐PrEP‐ared and AMPrEP using rank sum tests for continuous variables and Pearson’s Incidence rates (IRs) of HIV, HCV, syphilis, chlamydia and gonorrhoea per 100 person‐years were estimated by dividing the number of incident infections by person‐years of observation. We defined an incident HCV infection as the first newly diagnosed HCV infection (anti‐HCV positive and/or HCV RNA positive) among participants who were HCV‐negative at baseline (anti‐HCV negative and/or HCV RNA negative). Repeated HCV infections were not included in the analyses because Be‐PrEP‐ared only determined anti‐HCV and hence reinfections could not be identified. For infectious syphilis (stage 1 or 2), chlamydia and gonorrhoea repeated infections over time were included in the analysis. In case of an incident HIV or HCV infection, we assumed the infection occurred at midpoint between the last negative and the first positive test, after which time at‐risk stopped. All follow‐up time was considered time at risk for syphilis, chlamydia and gonorrhoea. IRs were estimated for the overall study population and stratified by PrEP regimen used at the moment of STI diagnosis. Incidence rate ratios (IRRs), adjusted for age and testing frequency per total months in follow‐up (as participants who test more frequently have a higher chance of getting a positive STI diagnosis), were estimated using Poisson regression with a random‐intercept to account for between‐city variability.All analyses were performed using Stata .3p = 0.004), less often had a STI in the six months prior , and less often had used PEP in the six months prior .We included 571 MSM in the analysis, 197 (34.5%) from Be‐PrEP‐ared and 374 from AMPrEP (65.5%). Median age was 39 years in both studies  participants chose edPrEP  never switched between PrEP regimens, 96 (17.2%) switched once and 81 (14.5%) more than once. In total, 228 switches were reported, of which 153 (53.1%) were from the daily to the edPrEP regimen. After 28 months, 23.5% of participants used edPrEP.3.2Two participants acquired HIV during follow‐up; both were using the daily PrEP regimen. One stopped taking PrEP , whereasAt baseline, 21 participants had evidence of current or past HCV infection . Seventeen new HCV infections were diagnosed during 993.9 person‐years of follow‐up . HCV IR did not differ between periods of daily and event‐driven use  . During 1018.9 person‐years, 116 new syphilis infections were diagnosed . Syphilis IR did not differ between periods of daily and event‐driven use .During follow‐up, 368 participants (65.9%) were diagnosed with one or more incident chlamydia or gonorrhoea infections. Overall IR of any chlamydia or gonorrhoea was 80.2 per 100 person‐years (95% CI 75.0 to 85.8). Incidence of any chlamydia or gonorrhoea was 61% higher during daily compared to event‐driven use    and 16% HIV incidence was low, consistent with previous PrEP studies among MSM . No seroThe following limitations should be taken into account. First, we only determined HCV‐antibodies in Be‐PrEP‐ared and hence could not include HCV reinfections in our analysis. In addition, although we included extra STI screening visits at the STI clinics between scheduled study visits, participants might have had additional STI visits outside of the study centres, so reported STI IRs may be underestimates. Second, the majority of participants were white and highly educated, thus results are not generalizable to the entire MSM population. Third, when both studies started, edPrEP was less well known, and not yet included in the WHO guideline . This ma4A quarter of participants chose edPrEP at baseline. Although switching between regimens was common, the proportion of participants using edPrEP was similar at 28 months. PrEP programmes should enable individuals to adapt their PrEP use to their needs (e.g. by providing more options than daily PrEP and giving information on how to safely switch regimens). The lower incidence of chlamydia and gonorrhoea among edPrEP users may suggest that less frequent STI testing of this group could be considered, even though the IRs of HCV and syphilis were similar. Further research on proportions and characteristics of edPrEP users and optimal STI testing frequency is warranted.The study medication for the Be‐PrEP‐ared study of Antwerp and the Amsterdam PrEP study was provided by Gilead Sciences based on unconditional grants. EH received advisory board fees from Gilead Sciences and speaker fees from Janssen‐Cilag, both paid to her institute. UD received unrestricted research grants and speaker’s fees from Gilead Sciences, paid to his institute. HJCdV received grants from Medigene, and advisory board and speaker fees from Gilead Sciences, Medigene, Abbvie, Janssen‐Cilag and Willpharma, paid to his institute. MP received unrestricted research grants and speaker’s fees from Gilead Sciences, Roche, Abbvie and MSD, paid to her institute. MSvdL served on an Advisory Board of MSD, paid to his institute. All other authors declare no competing interest.BV, TR, EH, MP, ML, MSvdL, HdV and UD conceptualized and designed the Tale of Two Cities project and obtained funding. ZY, VJ and MSvdL were involved in the data analysis. VJ, TR, ZY, MSvdL, UD, HZ, LC, MvdE, HdV, KW, TS, BV, MP, ML and EH were involved with interpretation of the data. VJ drafted the manuscript. All authors read and approved the final manuscript.The Be‐PrEP‐ared study (EudraCT 2015‐000054‐37) was approved by the institutional review board of the Institute of Tropical Medicine, Antwerp (988/15), and the ethics committee of the Antwerp University Hospital (15/25/255). The AMPrEP study obtained ethical approval from the ethics board of the Academic Medical Center, Amsterdam, the Netherlands (NL49504.018.14). AMPrEP was registered at the online Dutch trial registry (NTR5411). All participants provided written informed consent.


To quantitatively evaluate upper limb ataxia using a novel pen‐like sensor device in patients with spinocerebellar ataxia (SCA) and to assess its validity, reliability, and sensitivity to disease progression.We designed a cross‐sectional and longitudinal study of patients with SCA and healthy controls. Upper limb ataxia was evaluated using a device that measures the three‐dimensional position every 10 msec. Participants were instructed to move a pen‐like part of the device iteratively between two buttons. We evaluated the time, length, velocity, and variation coefficient of the stroke, and calculated the distortion index using the mean squared error. The following scales were also evaluated: Scale for the Assessment and Rating of Ataxia (SARA), the International Cooperative Ataxia Rating Scale (ICARS), and the nine‐hole pegboard test. Subjects were followed 12 months after the baseline evaluation.r = 0.647 and 0.722, respectively). Test–retest reliability was high for most of the ataxia indices. In the longitudinal analysis, the distortion index showed high standardized response mean and adjusted effect size, regardless of disease severity.A total of 42 patients with SCA and 33 healthy controls were enrolled and evaluated. For all ataxia indices measured using the device there were significant differences between healthy controls and patients with SCA. Among the ataxia indices, the distortion index showed the strongest correlation with the SARA and ICARS upper limb score (Pearson's Our study demonstrated that the distortion index is a reliable functional marker that is sensitive to longitudinal change in patients with SCA. Spinocerebellar ataxia (SCA) is a subset of hereditary progressive neurodegenerative disorders with ataxia, characterized by unsteadiness of gait, abnormal eye movements, slurred speech, and impaired coordination of limb movements.There is currently no effective treatment to slow the progression of SCA, although several clinical trials for potential disease‐modifying treatments, including cell therapy, are now underway.The present cross‐sectional and longitudinal study aimed to develop a quantitative evaluation method for upper limb ataxia for patients with SCA using a novel pen‐like sensor device and to evaluate the validity, reliability, and sensitivity to disease progression of the measured indices.This study was conducted according to the Declaration of Helsinki, the Ethics Guidelines for Human Genome/Gene Analysis Research, and the Ethical Guidelines for Medical and Health Research Involving Human Subjects endorsed by the Japanese government. The study protocol was approved by the ethics review committee of Nagoya University Graduate School of Medicine. All participants were informed of the purpose of the study and provided written informed consent.Patients who were clinically or genetically diagnosed with SCA were recruited. The principal inclusion criteria were as follows: (1) genetically confirmed SCA or (2) cerebellar ataxia, which was inherited as an autosomal dominant trait. Patients with secondary cerebellar ataxia caused by alcohol, drugs, infarction, autoimmune diseases, or paraneoplastic syndromes were excluded. Patients who had severe complications, including heart, or respiratory failure or cognitive dysfunction were excluded. Age‐and sex‐matched healthy controls (HCs) with no history of neurological abnormalities were also recruited. All participants were evaluated at Nagoya University Hospital from November 2017 to April 2020.We developed a measurement system that consisted of a commercially available device , which accurately measures a three‐dimensional position and applies force feedback on the uses' hand, and four buttons Fig. . This tiParticipants were instructed to sit in a chair in front of the device and move a pen‐like part between the two buttons 9.5 times as quickly as possible. They performed the same tasks along four different routes Fig.  in orderIn addition, we developed a distortion index, which is a measurement to assess decomposition and dysmetria. We approximated the measured trajectories with a smooth nonlinear function using three different regression models: B‐spline regression, restricted cubic spline regression, and polynomial regression. The deviation of the observed trajectory from the approximate curve was calculated using the mean squared error (MSE) and the MSE values obtained from multiple trajectories for each patient was averaged and defined as the distortion index .We evaluated the severity of ataxia using functional scales. The SARA score ranged from 0 (no ataxia) to 40 (severe ataxia), and the ICARS score ranged from 0 (no ataxia) to 100 (severe ataxia). The severity of upper limb ataxia was evaluated using the upper limb SARA score, which consisted of three items , with a range of 0–12, and the upper limb ICARS score, which consisted of five items , with a range of 0–36. Participants also performed the nine‐hole pegboard test (9HPT) using a Rolyan 9HPT apparatus and a plastic one‐piece model using both hands.Participants underwent motor functional assessments and evaluations using the device at each study visit. All participants were followed 12 months after the baseline evaluation to assess longitudinal changes. To evaluate test–retest reliability, consenting participants were assessed twice at an interval of 1–5 weeks.t‐tests were used for comparisons of variables between patients with SCA and HCs. Pearson's correlation coefficients were used to identify correlations between parameters. A p value <0.05 was considered significant, and correlation coefficients (r) > 0.6 were interpreted as strong.F‐value, which was set at 0.05. The equation was selected according to the highest multiple correlation coefficient. Test–retest reliability was assessed using intra‐class correlation coefficients (ICCs), specifically ICC . ICC values <0.5 indicated poor reliability, values between 0.5 and 0.75 indicated moderate reliability, values between 0.75 and 0.9 indicated good reliability, and values greater than 0.9 indicated excellent reliability.SDbaseline2+SDfollowup2/2 × 1−ICC. Minimal detectable changes (MDCs) at the 95% confidence level (MDC95) was calculated using the formula 1.96 × 2× SEM. MDCs are defined as a change beyond measurement error and helps to provide a better interpretation of change.2× 1−r, where r is Pearson's correlation coefficient.Statistical analyses were performed using the SPSS 25.0 J software  or SAS version 9.4 . All data are presented as means ± standard deviations (SD), unless otherwise stated. Chi‐squared tests and unpaired Student's Anonymized data of the current study will be made available to qualified investigators upon request.A total of 42 patients with SCA and 33 HCs were recruited and evaluated at baseline. A total of 12 patients and 17 HCs were lost to follow‐up. A final total of 30 patients with SCA and 16 HCs were analyzed after 12 months of follow‐up . AUC of the deviation to discriminate patients with SCA from HCs and parameter for fitting of the nonlinear model (B). SCA, spinocerebellar ataxia; MSE, mean squared error; AUC, area under the curve; HCs, healthy controls.Figure S4. Correlations between each functional measurement and ataxia index and disease duration, SARA total score and SARA upper limb score in patients with SCA. Correlations between nine‐hole peg test and disease severity‐related parameters: disease duration, SARA total score, and SARA upper limb score (A–C). Correlations between velocity and disease severity‐related parameters (D–F). Correlations between distortion index B‐spline and disease severity‐related parameters (G–I). Correlations between distortion index restricted cubic spline regression (RCS) and disease severity‐related parameters (J–L). Correlations between distortion index polynomial regression (POLY) and the disease severity‐related parameters (M–O). Significant correlation coefficients and p values are annotated. SCA, spinocerebellar ataxia; SARA, Scale for the Assessment and Rating of Ataxia.Table S1. Baseline characteristics between patients with SCA with and without follow‐up.Table S2. Ataxia indices in each hand of SCA and healthy subjects.Table S3. Reliability and minimal detectable changes for the movement indices of the device.Table S4. Longitudinal analyses of functional measurements and indices in SCA patients with and without non‐ataxia symptoms.Table S5. Longitudinal analyses of functional measurements and indices in SCA patients at early and late stages.Table 6. Longitudinal analyses of functional measurements and indices in SCA3 patients (n = 7).Table S7. Longitudinal analyses of functional measurements and indices in SCA6 patients (n = 6).Table S8. Longitudinal analyses of functional measurements and indices in SCA31 patients (n = 8).Supplemental methods. Methods for evaluation procedure, data preprocessing, calculation of the distortion index, and statistical analyses.Supplemental References. Reference for statistical analysis.Click here for additional data file.


The purpose of the current study was to compare the postoperative complications, overall survival and disease-free survival in young and old gastric cancer patients after gastrectomy using propensity score matching (PSM).Adult patients (aged ≥ 18 years) who underwent gastrectomy for gastric cancer in a single clinical center from January 2013 to December 2017 were enrolled continuously for retrospective analysis. To minimize the selection bias between the young and old groups, the PSM was conducted in this study.A total of 558 patients were included in this study, with 51 patients in the young group (aged ≤ 45 years) and 507 patients in the old group (aged > 45 years). After 1:1 matching according to PSM, 51 patients in the young group were matched to 51 patients in the old group. After PSM, there was no difference in the baseline information. In terms of short-term outcomes, no difference was found in operation time (P = 0.190), intraoperative blood loss (P = 0.336), retrieved lymph nodes (P = 0.948), blood transfusion (P = 0.339), postoperative hospital stay (P = 0.194), or postoperative complications (P = 0.477) between the two groups. For overall survival, no statistically significant difference was found in all stages (P = 0.383), stage I (P = 0.431), stage II (P = 0.875) or stage III (P = 0.446) gastric cancer. Furthermore, regarding disease-free survival, no differences were found between the two groups in all stages (P = 0.378), stage I (P = 0.431), stage II (P = 0.879) or stage III (P = 0.510) gastric cancer.Age might not be an independent prognostic factor for short-term outcomes, OS, or DFS in gastric cancer patients who underwent gastrectomy. The pTNM stage of GC might be an independent prognostic factor for OS and DFS. Gastric cancer (GC) is the fourth leading cause of cancer-related death in the world, and nearly 27,600 new cases were diagnosed in the USA in 2020 . DespiteIt is generally believed that the onset of GC mainly occurs in old patients . HoweverYoung patients were considered to have a poor prognosis compared with old patients in most studies. Lower differentiated histology and diffuse infiltration of the malignancy were found in young patients due to delayed detection , 9. HoweThus, the purpose of the current study was to compare postoperative complications, OS and disease-free survival (DFS) in young and old gastric cancer patients after gastrectomy using PSM.The medical records of 643 adult patients (aged ≥ 18 years) who underwent gastrectomy for gastric cancer in a single clinical database from January 2013 to December 2017 were enrolled continuously for retrospective analysis. The study was approved by the ethics committee of our institution , and all patients signed informed consent forms. This study was conducted in accordance with the World Medical Association Declaration of Helsinki as well.Patients who were diagnosed pathologically with GC and undergoing gastrectomy were included in this study (n = 643). The exclusion criteria were as follows: 1) palliative gastrectomy (n = 23); (2) remnants of gastric cancer (n = 12); (3) other malignant tumors presented synchronously (n = 9); and (4) incomplete medical records before gastrectomy (n = 41). The flowchart and inclusion and exclusion criteria are shown in Fig.  palliatiAccording to the 5th edition guidelines of the Japanese Gastric Cancer Association (JGCA) in 2018, gastrectomy with D2 lymph node dissection is the standard treatment for patients with resectable GC . The folThe patients were divided into two groups: patients aged ≤ 45 years were considered the young group, and patients aged > 45 years were considered the old group. The pathological stages of gastric cancer in this study were defined by consulting by the 5th edition guidelines of the JGCA published in 2018 . PostopePerioperative information and follow-up data were collected for analysis. The perioperative information included baseline information, operation time, intraoperative blood loss, retrieved lymph nodes, postoperative hospital stay, type of resection, reconstruction methods, pTNM stage and postoperative complications. Follow-up data, including OS and DFS, were collected as well.To minimize the selection bias between the two groups, PSM was conducted in this study , 16. NeaContinuous variables are expressed as the mean ± SD, and an independent-sample t test was used to compare the difference between the young and old groups. Frequency variables are expressed as n (%), and the Chi-square test or Fisher’s exact test was used. Cox regression analyses were performed to identify predictive factors for OS and DFS. Data were analyzed using SPSS (version 20.0) statistical software. A bilateral P value of < 0.05 was considered statistically significant.A total of 643 adult patients aged ≥ 18 years) who underwent gastrectomy were retrospectively analyzed in this study. According to the exclusion criteria, 558 patients were included in this study. There were 51 patients in the young group (aged ≤ 45 years) and 507 patients in the old group (aged > 45 years). Before PSM, there was a significant difference in the baseline characteristics, including sex (P = 0.000), comorbidities (P = 0.014), preoperative albumin levels (P = 0.000) and pTNM stage (P = 0.026), which are summarized in Table 8 years wAfter 1:1 matching according to PSM, 51 patients in the young group were matched to 51 patients in the old group. After PSM, there was no significant difference in any baseline characteristics between the two groups Table .Table 2BShort-term outcomes were compared between the two groups, and there were no differences found in operation time (P = 0.190), intraoperative blood loss P = 0.336), retrieved lymph nodes (P = 0.948), blood transfusion (P = 0.339), postoperative hospital stay (P = 0.194), or postoperative complications. The postoperative complications were graded by Clavien-Dindo classification, and no differences were found in overall complications (P = 0.477) or major complications (P = 1.000) between the young group and the old group  months. The comparison of OS was performed between the young group and the old group. No statistically significant difference was found between the two groups in any stage (P = 0.383)  stage of malignancy was related to OS (P = 0.008) and DFS (P = 0.009), and age was not a predictor in terms of OS (P = 0.531) or DFS (P = 0.572) Table .Table 4MA total of 558 patients, including 51 patients in the young group (aged ≤ 45 years) and 507 patients in the old group (aged > 45 years), were analyzed in this study. After 1:1 matching according to PSM, 51 patients in the young group were matched to 51 patients in the old group. After PSM, the baseline information of all the patients was not significantly different. In terms of postoperative complications, no difference was found in overall complications or major complications between the two groups. In addition, there were no significant differences in OS and DFS between the two groups.Age was considered an important factor correlated with the prognosis of malignant tumors in previous studies . SimilarIn addition, only a few studies have conducted PSM to control for selection bias , and it Postoperative complications tend to directly affect the prognosis of patients who underwent gastrectomy, and major complications might also affect OS , 23. It It is widely accepted that the OS of GC patients after gastrectomy is the major outcome indicator to determine the effectiveness of treatment . There iThe quality of life of patients with malignant tumors depends mostly on DFS, and the recurrence of tumors can result in difficult treatments and imminent death. Young patients might experience more cancer-related deaths, and the causes of death in old patients might be closely related to other comorbidities . NotablyThis current study had several strengths. First, this study was carried out using PSM to minimize the selection bias caused by baseline information, and subgroup analysis was conducted for each tumor stage to analyze OS and DFS. Second, previous PSM studies reported that age had a significant impact on survival; however, in this study, no significant difference was found.There were some limitations in the current study. First, this is a single-center retrospective study with a small sample size, which might cause bias. Second, to decrease the selection bias on baseline information, PSM analysis was conducted in this study; however, another bias existed due to the smaller sample size in the subgroup analysis of different stages. Third, the median follow-up time was relatively short, especially for the early stage of tumors. Thus, a larger sample size and multicenter prospective randomized controlled trials should be conducted in the future.In conclusion, age might not be an independent prognostic factor for short-term outcomes, OS, or DFS in gastric cancer patients who underwent gastrectomy. The pTNM stage of GC might be an independent prognostic factor for OS and DFS.


Bilateral cochlear implant (CI) implantation is increasingly used in the auditory rehabilitation of bilaterally deafened adults. However, after successful unilateral implantation, objective patient counselling is essential.We investigated the extra benefit of a second CI in adults in terms of health-related quality of life, tinnitus, stress, anxiety, depression, quality of hearing, and speech recognition. Hearing ability was assessed by using the Freiburg monosyllable speech discrimination test (FB MS) and the Oldenburg sentence test with azimuth variations. In a prospective patient cohort, we administered validated questionnaires before a CI, after a first CI and after a second CI implantation.The study included 29 patients, made up of nine women and 20 men. The median time between the first and the second implantation was 23 months. The mean total NCIQ score and TQ before a CI improved significantly after both implantations. Stress, anxiety, and depression were stable over time and were not significantly affected by CI implantations. Speech recognition with noise significantly improved after the first and again after the second CI. Correlation analysis showed a strong connection between auditory performance and HRQoL.We demonstrated that a unilateral CI benefitted many fields and that the second sequential CI leads again to additional improvement. Bilateral CI implantation should, therefore, be the standard form of auditory rehabilitation in deafened adults. The number of bilaterally deafened adults is increasing, mainly due to a constantly ageing society. A bilateral CI is a widely accepted medical practice in infants and children h . HoweverThe list of indications for cochlear implantation has expanded in the last decade. As a result, increased numbers of adult post-lingually deafened patients are scheduled for a second CI. To measure the impact of CIs on patients' lives, our study group uses a test battery, which in addition to the auditory performance, assesses the subjective perception of stress factors, patients' resources, ways of coping with stress, and the degree of anxiety and depressive symptoms . In the Between July 2010 and May 2016, 29 bilaterally deaf adult patients who were scheduled for a bilateral CI implantation were included in this study. The implantations were performed sequentially, and the median time between the implantations was 23 months. Patients were interviewed before the first implantation and six months after the first and the second implantation.The cohort consisted of nine women and 20 men, averaging 57.63 years of age. The period of deafness was determined based on the reported time when a patient felt that wearing a hearing aid had stopped being of significant benefit, and averaged 18.92 years. The mean value Freiburg monosyllable speech discrimination test (FB MS) for implanted ears was 6.05 (SD = 11.85), while on contralateral ears, it was 10.87 (SD = 16.42).For each patient, we collected data at three time points: prior to the first CI, six months after the first implantation, and again six months after second implantation. Patients filled paper questionnaires, as described by Bruggemann et al. (2017), including the Nijmegen Cochlear Implantation Questionnaire (NCIQ), the Tinnitus Questionnaire (TQ), the Perceived Stress Questionnaire (PSQ), the brief German COPE (COPE), the General Depression Scale (ADS), and the Generalised Anxiety Disorder (GAD) –22. SubjSpeech perception was measured in all patients before the CI, using the FB MS. This involved presenting patients with 2 × 20 monosyllabic words at a volume of 65 dB HL in quiet conditions. After the first and second CIs, the FB MS and Oldenburger Satz test (OLSA) were used . The OLSThe primary objective of this prospective analysis was to identify predictive values for a successful second CI, defined by an increase in health-related quality of life and better performance in the monosyllable test and the more difficult speech perception tests.p < 0.05.Twenty-nine patients were included in this study. Statistical evaluation was carried out using the software program SPSS . Frequency tables, mean values, and standard deviations were calculated descriptively. The Kolmogorov–Smirnov test was used to test for distribution normality, and the Wilcoxon test was used to test the significance of differences before and following implantation. Although some quantitative data were available, there were insufficient data for regression analysis, and so we report effect sizes herein, using Pearson's correlation test. Significance was set at p < 0.000) and then improved further to 63.28 ± 18.53 (p = 0.012) following the second CI . After the second CI, the FB MS improved further to 66.43 ± 19.57 (p = 0.002) with bilaterally active CI. After the second CI operation, both ears were also investigated with only one active CI. The second operated ear performed slightly better 56.59 ± 19.24 versus the first operated ear 50.68 ± 22.16. However, the difference was not statistically significant.The mean value FB MS for the first implanted ears was 6.05 ± 11.85; on contralateral ears, it was 10.87 ± 16.42. After first CI, FB MS increased to 48.69 ± 28.94 (p = 0.002). If the speech was presented to the first implanted ear, and noise to the deaf ear, the mean OLSA SNR was − 4.241 ± 5.338 dB respective to the contralateral CI − 3.518 ± 5.00 dB SPL. The difference, in this case, was not significant (p = 0.115). Speech understanding measures were justifiably the lowest observed when noise was presented to the CI ear and speech to the deaf ear at 7.231 ± 4.65 dB SPL in unilaterally implanted patients.When speech and noise were presented from the front, the mean OLSA SNR was 0.368 ± 3.109 dB SPL with a bilateral CI, while the mean OLSA SNR was 1.881 ± 4.459 dB SPL with a unilateral CI. The 1.513 dB SPL difference was significant  and decreased further after the second to 11.00 ± 13.44 (p = 0.041). Before any CI, five patients (23.8%) reported decompensated tinnitus . After the first CI, two patients in the decompensated tinnitus subgroup achieved intermediate levels . One patient reported a complete absence of tinnitus, and two reported continuous high tinnitus annoyance. These two patients profited from the second CI, with one achieving low-level and the other intermediate-level tinnitus. Each of the three patients (14%) with intermediate tinnitus annoyance before the first CI reported a decrease to low tinnitus levels  after the procedure, and this reduction persisted after the second CI. All subscales scores, including emotional and cognitive distress, intrusiveness, auditory perceptual difficulties, sleeping disturbances, but somatic complaints, declined significantly after the first CI and again after the second CI  reported tinnitus of some kind before CI implantation, with a mean total TQ score of 27.52 ± 19.67. TQ decreased after the first CI to 15.68 ± 20.37 (p = 0.037), and this value remained stable after the second procedure  were observed after the second implantation (see Table p = 0.546 and p = 0.862).The mean score of the ADS depression questionnaire was 13.28 ± 8.84 before any CI, and 10 of the 29 (34.4%) patients scored 16 or more, which is a degree of depression that is understood to be deserving clinical treatment. There was a slight reduction after the first CI, and a very slight increase after second CI. However, these changes were not significant (p = 0.049). After the second CI, it was 4.10 ± 3.90 (p = 0.807). Before CI, nine of the 29 (31%) patients reported at least mild anxiety (4–9 pt.) [The mean of the GAD-7 questionnaire, which assesses fear, was 3.75 ± 4.56 pre-CI, which indicates minimal to mild anxiety. After the first CI, the mean GAD-7 declined to 3.29 ± 3.94 (4–9 pt.) , and thiPearson correlations were performed in order to gain a better understanding of the relations between HRQoL (NCIQ), tinnitus distress (TQ), stress (PSQ), generalised anxiety disorder (GAD-7), depression (ADS), coping (COPE), subjective quality of hearing (OI), and hearing performance with background noise. The heat maps summarise the correlations after the first and after the second CI see Fig. . For theIn the present work, we used a prospective study to follow 29 post-lingually bilaterally deaf patients from the moment when they were scheduled for the first cochlear implantation, up to six months after the second cochlear implantation. At present, there is an ongoing global discussion about bilateral cochlear implantation as a standard procedure for bilaterally deafened people. Apart from medical and audiological issues, this discussion involves health politics. Here, we provide evidence about the medical, psychological, and social benefits in patients who were bilaterally, subsequently implanted with CI.One of our main findings was that HRQoL significantly improved, not only after the first CI implantation but also after the second CI. This remarkable additional improvement was seen in all subdomains of the NCIQ: primary sound perception, advanced sound perception, speech production, self-esteem, activity, and social interactions. Our results corroborate these of others , 12, 23 In addition to the quality of life, speech comprehension, and directional hearing improved after first and after second CI implantation, in agreement with already published data . Also, tMany studies report improved speech perception—and notably, directional listening—after a second CI , 24. TheEvidence accumulates about the reduction in tinnitus annoyance being closely related to the improvement in HRQoL –29. In tWe also demonstrated that the subjective ability of hearing correlates positively with the HRQoL and negatively with elevated scores in the depression assessment  patients reported at least mildly severe anxiety , which n. This fact produces an analytical heterogeneity. Additionally, the current study could not find predictive values by using regression analysis. The improvement in the HRQoL after the second CI might also be explained as a result of the better ear and upstream connections being suppressed over a shorter period, and therefore hearing performance and experience after the second CI being judged by CI recipients as very satisfactory.A great deal of effort has been spent on identifying the predictive values for the outcome of a second CI. In summary, no feature of an individual can predict the outcome—even advanced age, a long duration of deafness, or a long interval between implantations are not significant negative predictors . HoweverThe present study demonstrates that in bilaterally deafened adult patients, a second CI implantation improves HRQoL, tinnitus annoyance, speech comprehension under noisy conditions, and subjective quality of hearing compared to a one-sided CI condition. The extent of improvement in auditory performance was positively correlated with the improved health-related quality of life. No effect of implantation on depression, stress, and anxiety disorders was observed.There are no measures currently available after fitting the first CI that help predict whether a patient is a suitable candidate for a second CI; however, nearly all patients profit from a second CI in several respects. Specialists should thus actively talk with patients about a second CI in unilateral CI recipients to inform them objectively about the risks and possible benefits of the second implantation. A bilateral CI implantation should be encouraged in most cases as a standard form of auditory rehabilitation in deafened children and adults—with very few exceptions.


Microaggressions, a concept originally introduced for ethnic minorities, represent subtle day‐to‐day discrimination, damaging the psychological health and well‐being of lesbian, gay, bisexual, and transgender individuals as well. This study aimed to assess whether microaggressions occur in psychotherapists’ assessments of clients who identify as either lesbian or transgender woman when compared with those identifying as heterosexual woman.N = 5) and microaggressive (N = 5), used to determine a clinical impression of the client. A repeated measure analysis of variance was conducted to understand the likelihood of clients of different gender identity and sexual orientation receiving microaggressions.The study included a sample of 135 licensed psychotherapists (110 cisgender women and 25 cisgender men). Participants were presented with an audio file of a woman introducing herself during her first therapy session. Three versions were presented: a transgender, a lesbian, and a heterosexual client. Participants were asked to assess the clinical relevance of 10 questions defined as neutral (Participants were more prone to consider microaggressive questions relevant where the client identified as either lesbian or transgender.The findings highlighted a bias against lesbian and trans women in evaluating the relevance and usefulness of clinical information while making a psychological assessment of a case. In an eTransgender people also experience situations in which they necessarily have to come out , which can expose them to an increased risk of discrimination and its consequences  link of the patient's gender identity or sexual orientation to the problematic or symptomatic aspects presented. We also expect that the self‐assessed emotional arousal generated by a patient belonging to a marginalized group will moderate the relevance attributed to microaggressive questions for the transgender and lesbian condition. Moreover, starting from the evidence of the relation between emotional arousal and automatic responses, we hypothesized highly activated clinicians to be more prone to use microaggressions toward sexual and gender minorities than clinicians at moderate levels of arousal.2To test whether psychotherapists could enact microaggressions in the context of the first consultation with a client, we had to create ad hoc questions that were relevant to the presented case. For this purpose, we asked 84 early career psychologists to listen to the audio files, that were then used in the main study, and produce up to 10 questions they would have asked the client to have a better understanding of the clinical case.Participants generated 575 questions in total, with an average of 6.8 questions each. Three different raters (blind to the study's hypotheses) independently classified each of the items as nonmicroaggressive, possibly microaggressive, or microaggressive. The raters were trained to recognize microaggressions; they were provided with the definition of microaggressions several examples of microaggressions targeting sexual and gender minorities. Approximately 30.9% of the sample produced at least one microaggressive question (rated as microaggression by two out of three raters). Only the questions from the original pool of items on which all three raters agreed upon were used to construct the questions for the following study.33.13.1.1η2partial = 0.25) suggested a sample size of N = 120 for a 3 between‐participants × 2 within‐participants analysis of variance (ANOVA). In total, 135 licensed psychotherapists took part in the study, all of whom self‐identified as cisgender, White Italian citizens, with 110 (81.5%) identifying as female. Other demographic characteristics of the sample are summarized in Table An a priori power analysis was conducted to estimate the required sample size using G*Power  introducing herself and describing her problems (related to panic) during the first session with the participant  to 7 (extremely relevant). The questions reflected both microaggressive and neutral themes. The questions generated from the pilot study were used to build five microaggressive (Cronbach's α = 0.90) and five neutral questions (Cronbach's α = 0.71) for the current study  Hayes, . Finally44.1To compare the relevance of microaggressive and neutral questions depending on the sexual identity of the client, we computed a 3  × 2  ANOVA with the first‐factor as between‐participants and the second‐factor as within‐participants.F  = 119.36, p < 0.001, ηp2 = 0.47. Independent from the experimental condition, neutral questions were considered significantly more relevant  than microaggressive questions . The ANOVA did not yield a significant main effect of condition: F  = 1.76, p = 0.18, ηp2 = 0.03. However, in line with the hypothesis, a significant interaction effect between condition and type of questions emerged: F  = 14.56, p < 0.001, ηp2 = 0.18. As displayed in Figure M = 5.70, SD = 0.75), than in the transgender client condition , t(86) = 2.22, p = 0.03, d = 0.48, 95% confidence interval [CI] [0.05–0.90]. The analysis revealed no differences between cisgender lesbian  and transgender conditions, t(95) = 0.92, p = 0.36, and between cisgender straight and lesbian conditions, t(83) = 1.36, p = 0.18. The post hoc comparisons of the relevance of microaggressive questions showed no difference between the cisgender lesbian  and transgender conditions , t(95) = 1.68, p = 0.09. Crucially, microaggressive questions were rated as significantly less relevant for the cisgender straight  than for the cisgender lesbian client, t(83) = 2.26, p = 0.03, d = 0.49, 95% CI [0.06–0.93], and the transgender client, t(86) = 3.94, p < 0.001, d = 0.85, 95% CI [0.41–1.29].The analysis suggested a main effect for the type of questions: F , 132 = 14.2α = 0.78), suggesting that both positive and negative items of the scale measured the same construct. A more theoretical reason is based on the argument that the positive affect and negative affect subscales of the PANAS both measure the latent construct of emotional arousal . At a high level of arousal ( = 2.55), therapists rated microaggressive questions more relevant for cisgender lesbian and transgender women than for cisgender straight women . At low levels of arousal ( = 1.85), the difference between cisgender straight and cisgender lesbian/transgender conditions was not significant .To ascertain whether emotional arousal influenced the relevance attributed to microaggressive questions, a moderation model was carried out through PROCESS  Hayes, . In this4.3F  = 0.001, p = 0.97, or any interaction effects with gender . The same results were obtained introducing sexual orientation as an independent variable . We also controlled for the effect of the therapists’ theoretical background  and years of practice on relevance evaluations. No interaction or main effects of theoretical background were identified , or for years of practice .To ascertain whether the therapists’ individual differences influenced their responses, we controlled for the effect of the therapists’ gender and sexual orientation. The first ANOVA yielded no significant main effect for gender, 5The aim of this study was to investigate the presence of microaggressive responses when psychotherapists work with clients of a marginalized sexual orientation or gender identity . The findings highlighted a bias in evaluating the relevance and usefulness of information while making a psychological assessment of a case. Data showed the effect of the client's gender identity and sexual orientation in rating the relevance of microaggressive questions by the therapists. Such questions were considered less important for the normative straight cisgender client and more important for the less conforming clients (cisgender lesbian and transgender). Conversely, asking general questions about symptoms, life events, and mental health history was considered less relevant for the assessment in the transgender client condition compared to the cisgender straight client condition.Results indicated that both cisgender lesbian and transgender women were equally targets of microaggressions. In addition, neutral questions were considered significantly less relevant only in the transgender condition compared with the cisgender straight condition. This result could be explained by considering the long history of pathologization of transgender identities. Once a client is labeled as transgender, they are immediately “diagnosed” by some of the psychotherapists that probably still rely on a pathologizing understanding of transgender identities. Thus, the main issues and complaints they bring to the consultation fade into the background as the therapist no longer needs additional information. Such bias relies on specific assumptions based on cisgenderism and heteronormativity .6In spite of these limitations, the study presents numerous strengths, including investigating microaggressions directed toward gender/sexual minorities “in vivo,” with a sample of licensed and experienced psychotherapists. This approach, rather than relying on the reported experiences of LGBT patients in therapy, allows observing directly the therapists' response to the client's disclosure of information. The use of an audio file in which the client introduces herself identifying as lesbian, trans, or heterosexual, allows the understanding of the intrinsic and implicit functioning of microaggressions through the approach of the therapist. Furthermore, the inclusion of emotional arousal among the study variables allows spreading lights on other dimensions that can potentially be targeted in the training of mental health professionals on LGBT issues.https://publons.com/publon/10.1002/jclp.23140The peer review history for this article is available at