How often do Medicare beneficiaries have an ambulatory follow-up visit after discharge from the emergency department, and is ambulatory follow-up associated with postdischarge outcomes?In this cohort study of 9 470 626 emergency department discharges from 2011 to 2016, most patients had ambulatory follow-up within 30 days, with lower rates among Medicaid-eligible beneficiaries, Black beneficiaries, and those treated at rural emergency departments. Ambulatory follow-up was associated with a higher risk of subsequent hospitalization but a lower risk of 30-day mortality.The findings of this study suggest that access to ambulatory care may be a key driver of outcomes among Medicare beneficiaries discharged from the emergency department. This cohort study examines the frequency and variation in ambulatory follow-up among Medicare beneficiaries discharged from US emergency departments (EDs) and the association between ambulatory follow-up and postdischarge outcomes. Ambulatory follow-up care is frequently recommended after an emergency department (ED) visit. However, the frequency with which follow-up actually occurs and the degree to which follow-up is associated with postdischarge outcomes is unknown.To examine the frequency and variation in ambulatory follow-up among Medicare beneficiaries discharged from US EDs and the association between ambulatory follow-up and postdischarge outcomes.This cohort study of 9 470 626 ED visits to 4728 US EDs among Medicare beneficiaries aged 65 and older from 2011 to 2016 who survived the ED visit and were discharged to home used Kaplan-Meier curves and proportional hazards regression. Data analysis was conducted from December 2019 to July 2020.Ambulatory follow-up after discharge from the ED.Postdischarge mortality, subsequent ED visit, or inpatient hospitalization within 30 days of an index ED visit.P < .001), Black race , and treatment at a rural ED  in the multivariable regression model. Ambulatory follow-up was associated with lower risk of postdischarge mortality  but higher risk of subsequent inpatient hospitalization  and ED visits , adjusting for visit diagnosis, patient demographic characteristics, and chronic conditions.The study sample consisted of 9 470 626 index outpatient ED visits to 4684 EDs; most visits (5 776 501 [61.0%]) were among women, and the mean (SD) age of patients was 77.3 (8.4) years. In this sample, the cumulative incidence of ambulatory follow-up was 40.5% (3 822 133 patients) at 7 days and 70.8% (6 662 525 patients) at 30 days, after accounting for censoring and for mortality as a competing risk. Characteristics associated with lower rates of ambulatory follow-up included beneficiary Medicaid eligibility (hazard ratio [HR], 0.77; 95% CI, 0.77-0.78; In this cohort study of Medicare beneficiaries discharged from the ED, nearly 30% lacked ambulatory follow-up at 30 days, with variation in follow-up rates by patient and hospital characteristics. Having an ambulatory follow-up visit was associated with higher risk of subsequent hospitalization but lower risk of mortality. Ambulatory care access may be an important driver of clinical outcomes after an ED visit. First, how often do Medicare beneficiaries discharged from the ED have an ambulatory follow-up visit within 7 and 30 days? Second, are certain patient or hospital characteristics associated with higher rates of ambulatory follow-up after ED discharge? Third, is ambulatory follow-up after ED discharge associated with differences in the postdischarge outcomes of mortality and subsequent ED or inpatient utilization?21 We further excluded visits resulting in discharge to a nursing or rehabilitation facility (680 064). We adhered to all Strengthening the Reporting of Observational Studies in Epidemiology (STROBE) reporting guideline for observational studies.22 The Harvard T.H. Chan School of Public Health institutional review board approved this study. Informed consent was waived because of the retrospective nature of the data.We examined a random 20% of Medicare beneficiaries from 2011 to 2016 who had an outpatient ED visit and were discharged home , Medicaid enrollment, and death date were obtained from the Medicare denominator file.24 ED visits were identified from the Medicare inpatient and outpatient claims files. We also examined hospital size, teaching status, urban or rural location, ownership, and safety-net status from linked 2014 American Hospital Association annual survey and Medicare Impact Files , or inpatient hospitalization within 30 days of ED discharge. For these outcomes, time to ambulatory follow-up was treated as a time-varying covariate, ensuring that only those follow-up visits occurring before the respective outcomes were considered when assessing the association between follow-up and the risk of these outcomes. Once the follow-up visit occurred, that patient’s hazard of the outcome occurring changed to the degree indicated by the hazard ratio (HR) from the model. This change is relative to other patients (at the same time after their ED visits) who have not yet had a follow-up visit.P < .05. Analyses were performed in SAS version 9.4 (SAS Institute).A Kaplan-Meier curve was generated to estimate the time-dependent probability of ambulatory follow-up after ED discharge, treating death as a competing risk and 30 days as the censoring time. To examine the associations between patient and hospital characteristics and ambulatory follow-up, we initially compared patient characteristics  and hospital characteristics  between patients who did and did not have an ambulatory visit within 30 days. These comparisons were formalized in a proportional hazards regression model with time until ambulatory follow-up as the outcome, death as a competing risk, and 30 days as the censoring time. We included patient and hospital characteristics, visit year, principal visit diagnosis category, and beneficiary chronic conditions to adjust for differences in severity between patients that may be associated with timing of follow-up. Finally, to account for clustering of patients within hospitals, we used the analogue of generalized estimating equations to adjust for correlation between the within-hospital residuals using the sandwich estimate of the covariance matrix. To avoid multiple testing, results are presented as HRs and 95% CIs. All hypothesis tests were 2-sided, and statistical significance was set at 29 We incorporated the year of visit, principal visit diagnosis, beneficiary demographic characteristics, and beneficiary chronic conditions as covariates and accounted for hospital-level clustering by stratifying by hospital. Stratification allowed each hospital to have its own unique baseline hazard for ambulatory follow-up, much like fixed effects for hospitals in linear and logistic regression models, with the added benefit that the hazards were not assumed to be proportional between hospitals. All patients within the same hospital shared the same baseline hazard so that patients with ambulatory follow-up were compared only with patients without ambulatory follow-up in the same hospital. For the outcomes of subsequent ED visits and inpatient hospitalizations, we incorporated mortality as a competing risk. For the association of ambulatory follow-up with each of these 3 outcomes, we show P values, HRs, and confidence intervals. Because of multiple testing, P < .0167 denotes significance.To examine the association between ambulatory follow-up and postdischarge outcomes, we specified 3 separate Cox regression models for the outcomes of mortality, all subsequent ED visits, and inpatient hospitalizations after discharge from the index ED visit. For each model, our primary exposure was ambulatory follow-up as a time-varying covariate to avoid immortality bias .We repeated our main models limiting follow-up to 7 days to check the robustness of the choice of censoring time as well as to qualitatively evaluate the proportional hazards assumption by checking whether the HR remained similar in early follow-up as in late follow-up. To assess potential differences by clinical condition, we repeated our main analyses separately for 10 clinical conditions with the largest magnitude risk of postdischarge mortality, defined by the regression coefficient for the outcome of mortality in our main Cox regression model. We also examined whether any patient-level associations between ambulatory follow-up and postdischarge events differed by hospital-level practice patterns and ambulatory care access. We first calculated hospital-level risk-adjusted rates of ambulatory follow-up using a linear probability model with follow-up rates as the outcome and adjusting for hospital random effects, principal diagnosis, and beneficiary characteristics. We then created 3 groups of hospitals based on their adjusted rates of ambulatory follow-up after ED discharge into high follow-up (top quartile), medium follow-up (middle 50%), and low follow-up hospitals (bottom quartile). We repeated our visit-level Cox regression models with ambulatory follow-up as a time-varying covariate as the primary exposure for each of the 3 postdischarge outcomes. We ran these models separately for high, medium, and low follow-up hospitals. Additionally, to examine whether rates of ambulatory follow-up were associated with other indicators of quality, we calculated the adjusted rates of each postdischarge outcome as well as the mean Hospital Compare overall star rating and proportion of patients giving a high rating of the Hospital Consumer Assessment of Healthcare Providers and Systems patient experience survey for each stratum of hospital follow-up.P < .001) as well as Black race . Treatment at a rural ED was associated with a 25% lower hazard of follow-up  after accounting for patient and hospital characteristics and chronic conditions.The study sample consisted of 9 470 626 index outpatient ED visits to 4728 EDs; most patients (61.0%) were women , and theP < .001) (P < .001) and a 22% higher hazard of an inpatient stay . The associations between all covariates in the model and risk of postdischarge outcomes are presented in eTable 2 in the Our initial sample of 9 470 626 ED visits included 3 711 826 beneficiaries, of whom 127 412 died and 42 668 were censored for termination of enrollment during the study period. Among the sample of 6 662 525 ED visits ending in discharge, there were 1 640 598 return visits to the ED and 885 761 subsequent hospitalizations. The 30-day cumulative incidence for postdischarge outcomes was 1.4% for mortality, 17.4% for postdischarge ED visit, and 9.4% for postdischarge hospitalizations when accounting for censoring and mortality as a competing risk . Ambulat < .001) . AmbulatP < .001) at 7 days compared with a 1% higher risk at 30 days. When we repeated our analysis separately for 10 high-risk conditions with the greatest risk of postdischarge mortality, our findings were similar  hospital-level rate of 30-day ambulatory follow-up was 65.1% (13.1%). Mean follow-up rates among hospitals in the bottom quartile, middle 50%, and highest quartile of ambulatory follow-up were 46.4%, 68.7%, and 76.9%, respectively. After repeating our analysis examining the association between ambulatory follow-up as a time-varying covariate and hazard of postdischarge events, there was again a consistent association between ambulatory follow-up and lower adjusted mortality. However, the magnitude of this association was greatest at hospitals with the highest follow-up rates  compared with medium  and low  follow-up hospitals . High foIn this study of 9 470 626 ED visits from 2011 to 2016, we found that approximately 29% of Medicare beneficiaries aged 65 years and older did not have an ambulatory follow-up visit within 30 days of ED discharge, while nearly 60% lacked follow-up with 1 week. Beneficiary Medicaid eligibility, Black race, and treatment at a rural hospital were associated with a lower rates of follow-up after ED discharge. Ambulatory follow-up was associated with approximately half the risk of 30-day postdischarge mortality but a 22% higher hazard of hospitalization. This association was present across individual conditions and hospitals but was strongest for those hospitals with the highest rates of follow-up after ED discharge, which tended to have better postdischarge outcomes in general. Additionally, these associations were strongest for conditions with the greatest risk of death after ED discharge in our sample.4 Within this high-risk population, our analyses showed that vulnerable subgroups, such as those who were Medicaid eligible, had even longer times to follow-up, as did Black beneficiaries, for whom disparities in health care access and quality have been well-documented.30 Lower mortality among those with ambulatory follow-up suggests that access to follow-up care may be a key driver of postdischarge outcomes and a potential target for reducing health care disparities.4Inconsistent rates of ambulatory follow-up after ED discharge among Medicare beneficiaries suggest an opportunity to improve care. Given the older age and greater chronic disease burden among Medicare beneficiaries relative to other ED patients, the immediate postdischarge period carries particularly high risk.32 Beneficiaries with follow-up visits had a higher risk of a downstream inpatient hospitalization and lower mortality compared with those without follow-up. Clinicians caring for Medicare beneficiaries after ED discharge may be appropriately referring persistently ill patients back to the hospital to prevent further deterioration. Rather than representing failed transitions of care, this subset of readmissions may represent follow-up clinicians serving as a safety net in this postdischarge period. Similar phenomena have been suggested for patients discharged after an inpatient hospitalization.31 Taken together, these findings suggest that while avoiding an ED visit or hospitalization is desirable, a subset of acute visits may actually prevent mortality; thus, policies with a disproportionate focus on reducing acute care utilization could lead to unintended harm.Our results also underscore the potential trade-off between reducing acute care utilization and improving quality by reducing mortality.35 While insurance type is an important determinant of access, we observed disproportionately lower rates of follow-up among non-White and Medicaid-eligible beneficiaries, even among traditional Medicare beneficiaries.17 Several factors likely contribute to these disparities, including racial discrimination in scheduling appointments, barriers to accessing nonemergency transport, and limited Medicaid expansion, less generous benefits, and a higher threshold for eligibility in some rural states.38 Among hospital characteristics, treatment at a rural ED has been shown to be associated with higher mortality rates among Medicare beneficiaries. Our findings are consistent with literature suggesting that barriers to ambulatory care access may be a key driver of poor health outcomes for Medicare beneficiaries treated in rural settings.40 These findings also serve as further evidence that hospitals serving populations in areas with disproportionately fewer resources may be penalized for factors beyond their control.41Our results are consistent with numerous studies documenting socioeconomic disparities in health care access and outcomes.42 However, Medicare beneficiaries account for growing share of ED visits, and lessons may be generalizable given that 60% of US residents have at least 1 chronic condition.44 Our study is observational, and therefore, the associations suggest, but do not prove, the benefit of ambulatory follow-up after ED discharge. The observational nature of our analysis and inability to completely control for patients who died before completing follow-up should be considered in future policy recommendations. We considered whether the association between higher rates of ambulatory care follow-up and lower postdischarge mortality was confounded by healthier patients who have fewer barriers to follow-up by virtue of their better health and thus are also more likely to survive. However, it is also possible that the bias goes the other way and that sicker patients are more likely to seek postdischarge care, leading to an underestimate of the benefit of ambulatory follow-up. Either a randomized clinical trial or daily monitoring of patient severity would be required to avoid or totally control for confounding at the time of the ambulatory visit. In this study, we controlled for patient characteristics, diagnosis categories, and chronic conditions to try to reduce confounding. The magnitude of the adjusted association seems too large to be entirely explained by unmeasured differences in patient severity alone. The fact that patients who had ambulatory follow-up were also more likely to have a subsequent hospitalization argues against this hypothesis, given that healthier patients generally do not use more hospital care. Furthermore, our findings were similar among hospitals with low rates of follow-up, where patients are more likely to experience barriers to accessing care relative to hospitals with high follow-up rates.Our study has a number of limitations. Findings among Medicare beneficiaries aged 65 years and older may not be generalizable to younger, healthier individuals with lower risk for mortality or readmission.In this study of Medicare beneficiaries aged 65 years and older who were discharged from US EDs from 2011 to 2016, nearly one-third (29%) lacked an ambulatory follow-up visit within 30 days. There was substantial variation among patients and hospitals in rates of follow-up after ED discharge, with lower follow-up rates among Medicaid-eligible beneficiaries, Black beneficiaries, and those treated in rural EDs. Ambulatory follow-up was associated with lower risk of postdischarge mortality but higher risk of a subsequent inpatient hospitalization. This association was observed for hospitals with high, medium, and low rates of follow-up but was greatest among hospitals with the highest rates of ambulatory care access. These findings suggest return visits to the ED or inpatient setting may be life-saving for some patients and that excess reductions in postdischarge acute care utilization may contribute to avoidable harm. Ambulatory follow-up care after ED discharge may be an important driver of outcomes among Medicare beneficiaries discharged from the ED.


During the intervention, MMT scores and AIS motor scores increased significantly . Improvements were stable in follow-up. AIS sensory scores, MAS, and SCIM were not modified significantly. MMT score prior to intervention was a good predictor of changes in walking speed  is dependent on the extent of remaining neural transmission in the corticospinal pathway. The aim of this proof-of-concept pilot study was to explore the effects of long-term paired associative stimulation (PAS) on leg muscle strength and walking in people with SCI. Five individuals with traumatic incomplete chronic tetraplegia  with no contraindications to transcranial magnetic stimulation (TMS) received PAS to one or both legs for 2 months . The participants were evaluated with the Manual Muscle Test (MMT), AIS motor and sensory examination, Modified Asworth Scale (MAS), and the Spinal Cord Independence Measure (SCIM) prior to the intervention, after 1 and 2 months of PAS, and after a 1-month follow-up. The study was registered at  Improving mobility in individuals with chronic spinal cord injury (SCI) remains a major clinical challenge , 2. RecoIt is important to develop non-invasive neuromodulation for restoration of locomotor function , 3, 5, 6Long-term application of PAS with the development of novel stimulation protocols has shown promising results in rehabilitation after SCI –15. ThisSeveral studies reported changes in the excitability of corticospinal projections to the lower limb muscles by means of PAS –21. RepoMuscle strength is strongly correlated to functional gait outcomes , 23. Weaclinicaltrials.gov (NCT03459885). Inclusion criteria were cervical incomplete SCI of traumatic origin and age 18–70 years. Exclusion criteria were contraindications for TMS or Magnetic Resonance Imaging (MRI), including the presence of intracranial metal objects, implanted electronics, epilepsy, pregnancy, high intracranial pressure, and any brain pathology visible in computer tomography or MRI  with chronic incomplete tetraplegia of traumatic origin  were enry or MRI . Initialy or MRI . Conventy or MRI . To ensuPAS was given for 2 months, 5 times per week during first 2 weeks (10 sessions) and 3 times per week during the subsequent interventions  with a figure-of-eight coil (outer diameter 70 mm). The patient's structural T1-weighted MRI  was employed in NBS System 4.3 to build an individual 3D head model. Gyral anatomy was utilized to identify a TMS hotspot, a site within the contralateral motor cortex where stimulation most readily elicited MEPs from the muscle of interest. The hotspots were defined for the  abductor hallucis or soleus muscles , tibialis anterior or extensor digitorum brevis muscles , vastus medialis or lateralis muscles , and gluteus maximus  using a built-in Eximia EMG device and surface electrodes . If no MEPs were obtained from the first-choice muscle, the second-choice muscle was used. The stimulated hotspots in each participant corresponded to stimulated nerves ; e.g., iThe Dantec Keypoint electroneuromyography device  and surface electrodes utilized for MEP recordings were used to determine mean latency, amplitude, and persistence of F-waves. For this purpose, 10 responses to 0.2-ms pulses at supramaximal intensity were recorded from the same muscles as in MEP measurements in each subject. PNS intensity used for PAS was defined for each participant individually as the minimum intensity inducing F-responses to single 1-ms pulses , 16.During PAS, 100-Hz PNS trains   were delIn all PAS sessions, TMS intensity was 100% of stimulator output (SO) and PNS intensity was defined individually for each peripheral nerve. The conduction times for lower motor neurons and upper motor neurons were estimated individually for each hotspot-nerve pair by measuring the minimum F latency and average MEP latency. ISI between TMS and PNS were calculated using the formula F-latency minus MEP latency  to coincAll patients tolerated TMS well. All patients were offered local skin anesthesia by EMLA ointment to reduce possible uncomfortable sensations caused by PNS , 28. ThrAn experienced physiotherapist specialized in SCI assessed the strength of gluteus maximus and medius, iliopsoas, quadriceps femoris, tibialis anterior, tibialis posterior, peroneus, long toe extensors, hamstrings, gastrocnemius, and hip adductors with the Manual Muscle Test (MMT). Spasticity was assessed with the Modified Ashworth Scale (MAS) from the hip , knee , and ankle (plantar flexors) muscles. Motor function was also evaluated with the standard American Spinal Injuries Association Impairment Scale (AIS). The physiotherapist did not participate in PAS and was blinded to the rule of nerve selection. The physiotherapist also measured gait function. For this purpose, the maximum distance that patients were able to walk without a break and without assistance of another person and the corresponding total time were measured and used to calculate mean walking speed. This test was selected to evaluate the functional outcome of PAS. MMT, AIS examination, and walking test were performed at each evaluation. A physician examined the patients' sensory function with AIS. The MAS, AIS examination of sensory function, and Spinal Cord Independence Measure (SCIM) were performed at the beginning of the intervention and at follow-up.The mean MMT score of all evaluated muscles in both legs and the mean MMT scores of the muscles innervated by each of the stimulated nerves  in both legs were calculated in each participant for each evaluation separately. The muscles with a value of 5 before the intervention (corresponding to full muscle strength) were excluded from the analysis as no further improvement could be detected by MMT. The differences between MMT scores obtained in Mid- and Pre-PAS (change during the first month of intervention), Post- and Pre-PAS (change during the entire intervention), and Follow-up and Pre-PAS (change during the intervention and the follow-up) were also calculated. The mean AIS motor scores in both legs were computed in each participant for each evaluation separately, excluding the muscles with a value of 5 before the intervention. Differences between mean AIS scores were computed as for calculations for MMT scores. Differences in walking speed in pre-PAS and post-PAS and in pre-PAS and follow-up were computed. Due to technical failure, the walking test was not performed for patient 1 in mid-PAS and the results of the second walking evaluation were not included in the group analysis. The sum of points obtained with the MAS was calculated for each of the four evaluations separately. The sum of AIS sensory scores and SCIM was computed for the pre-PAS and the follow-up.Statistical comparisons were performed in IBM SPSS Statistics  using the Friedman test and Wilcoxon signed-rank test where appropriate. Results are reported as medians. In addition, mean and standard error of the mean are shown in the tables and quartiles in figures. Regression analysis was used to assess the extent to which the total variation in the changes of walking speed (dependent variable) in the follow-up can be explained by the MMT score in the pre-PAS (independent variable). Three regressions were computed separately for all MMT scores, MMT scores of key muscles, and MMT scores of remaining muscles. Key muscles included gluteus maximus and gluteus medius, knee flexors, knee extensors, and ankle plantar flexors. This selection was based on previous reports on their significance particularly for walking speed improvement  and neurn = 5, χ2(3) = 10.563; p = 0.014, Friedman's test]. Post-hoc analysis with Wilcoxon signed-rank test revealed an increase of total MMT scores in Mid-PAS , Post-PAS , and follow-up  when compared with Pre-PAS. Partial MMT scores were increased by the PAS from Md = 2.40–3.67 . Post-hoc analysis with Wilcoxon signed-rank test revealed an increase of partial MMT scores in Mid-PAS , Post-PAS  and follow-up  when compared with Pre-PAS. Raw MMT scores are presented in Total MMT scores were increased by the PAS from Md = 2.58–3.58 . Post-hoc analysis with Wilcoxon signed-rank test revealed an increase of AIS motor scores in Post-PAS  and in follow-up  when compared with Pre-PAS. The differences between Mid-PAS vs. Pre-PAS  were not statistically significant. Raw AIS motor scores are presented in The median AIS motor scores increased from 3.00 (pre-PAS) to 4.33 (Follow-up) . Nevertheless, walking speed increased in the post-PAS in all ambulatory participants . Walking speed increased post-PAS in participants 1, 2, 4, and 5 . Participant 3, who was non-ambulatory before the intervention, could take several steps with an Eva support walker but this result was not quantifiable. Walking distance increased in 3 out of 4 ambulatory participants  = 18.564, p = 0.024, r = 0,872, p = 0.054], with an R2 of 0.861. The patients' predicted walking speed was equal to −0.187 + 0.005x  m/s. The adjusted y = −0.224 + 0.010x (MMT score of key muscles) was observed for the key muscle MMT score  with an R2 of 0.972 . Results of regression analysis with MMT scores of the remaining muscles were not significant .A significant regression equation was found for all MMT scores .No significant changes in spasticity as assessed with the MAS were observed in the mid-PAS . SCIM scores collected in the pre-PAS and the follow-up did not differ . Participants 4 and 5 improved their SCIM score by 15 and 9 points, respectively, mostly in Mobility subscales  reported slightly increased spasticity in his right hand and in the left leg as well as tiredness on week 2 of the intervention, which later disappeared. In week 5 he reported pain in his lower back while standing up or sitting down. The participant temporarily increased pain medication and received treatment from a physiatrist. The participant did not discontinue treatment. Other participants did not report any adverse effects.The current study provides the first evidence that long-term PAS may increase leg muscle strength. We observed that long-term PAS applied for 2 months increased the total lower limb MMT score by ~1 point in each muscle. This effect was stable after a 1-month follow-up. The results of AIS motor score revealed a similar increase of muscle strength by ~1 point per muscle. Quantitatively similar findings were obtained in previous studies , 29, wheIn the present study, PAS was applied in a heterogenous group of participants. Initial lower limb muscle strength measured by MMT score and AIS motor score varied between participants. Stable improvements in lower limb muscle strength were observed in all participants. This increases the external validity of the present study. These results support efficacy of long-term PAS and also demonstrate an increase of walking speed post-PAS. Walking distance increased in 3 out of 4 ambulatory participants. The design of the study, which investigated both lower limb strength and locomotion, limited the possibility of stimulation of only one limb in all participants for controlling the effects of stimulation. However, in Participant 1, only the stimulated leg improved .All participants had a standard rehabilitation schedule before the intervention and this schedule continued unchanged during the intervention and the follow-up period. The average changes of AIS motor scores during PAS ranged from 1.2 (in Mid-PAS) to 1.4 (in Post-PAS) points and were stable during follow-up. In SCI, motor improvements occur within the first 6–9 months and a plateau is usually reached by 12 months post-injury with minimal chance of further improvement . This exWe selected the nerves innervating the weakest muscles for stimulation. This approach also led to improvements in the muscles that were not directly innervated by the stimulated nerves. Improvements in weak muscles plausibly affected the entire limb by making walking easier and thus promoting increased use of all muscles. Activation of the neighboring regions of the motor cortex and peripheral nerves is also plausible.PAS was combined with a slight motor activation. Paired corticospinal-motoneuronal stimulation combined with a small level of isometric muscle contraction has been shown to significantly improve corticospinal transmission in people with SCI when compared with corresponding stimulation applied at rest . It is pA previous study revealed that high-frequency repetitive TMS applied immediately before robotic training sessions improved motor function in a chronic SCI patient . The autCorrelation and significant linear relations between initial MMT scores collected in pre-PAS and changes of walking speed after PAS indicate that MMT scores have predictive value for changes in walking speed after long-term PAS. Key muscles can provide even better prediction. Changes in MMT scores in these muscles were shown to predict changes in walking speed after locomotor training . RegressAge reduces neuroplasticity and the subsequent restoration of motor function after SCI , indepenIn our previous studies on PAS therapeutic effects in the upper limbs , 17, 29,The 80–120-min duration of one PAS session depended on the number of stimulated nerves. The protocol was generally well-tolerated by the patients. Considering the variability of available resources in different countries, further research is needed to evaluate whether PAS protocols that employ higher TMS frequency with shorter duration , 36 coulA limitation of this study was the small sample size; the results of this case series should be considered exploratory. The observed motor changes and functional outcomes following long-term application of PAS in lower limbs enhance our understanding of the therapeutic mechanism of PAS and contribute to further development of long-term PAS therapy. The results of this proof-of-concept pilot study indicate a possible beneficial effect of long-term PAS in the rehabilitation of traumatic tetraplegia.The raw data supporting the conclusions of this article will be made available by the authors, without undue reservation, to any qualified researcher.The studies involving human participants were reviewed and approved by The Ethics Committee of the Hospital District of Helsinki and Uusimaa. The participants provided their written informed consent to participate in this study.AR, SS, EK, and AS data acquisition. AR, SS, EK, JM, and AS research design, analysis, and interpretation of data, drafting the work and revising it critically for important intellectual content, final approval of the version to be published, agreement to be accountable for all aspects of the work.JM reports receiving travel expenses for lectures from Nexstim Inc outside the submitted work. The remaining authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


Socioeconomic inequalities in smoking cessation have led to development of interventions that are specifically tailored for smokers from disadvantaged groups. We aimed to assess whether the effectiveness of interventions for disadvantaged groups is moderated by tailoring for socioeconomic position.For this systematic review and meta-regression, we searched MEDLINE, PsycINFO, Embase, Cochrane Central Register, and Tobacco Addiction Register of Clinical Trials and the IC-SMOKE database from their inception until Aug 18, 2019, for randomised controlled trials of socioeconomic-position-tailored or non-socioeconomic-position-tailored individual-level behavioural interventions for smoking cessation at 6 months or longer of follow-up in disadvantaged groups. Studies measured socioeconomic position via income, eligibility for government financial assistance, occupation, and housing. Studies were excluded if they were delivered at the community or population level, did not report differential effects by socioeconomic position, did not report smoking cessation outcomes from 6 months or longer after the start of the intervention, were delivered at a group level, or provided pharmacotherapy with standard behavioural support compared with behavioural support alone. Individual patient-level data were extracted from published reports and from contacting study authors. Random-effects meta-analyses and mixed-effects meta-regression analyses were done to assess associations between tailoring of the intervention and effectiveness. Meta-analysis outcomes were summarised as risk ratios (RR). Certainty of evidence was assessed within each study using the Cochrane risk-of-bias tool version 2 and the grading of recommendations assessment, development, and evaluation approach. The study is registered with PROSPERO, CRD42018103008.2=54·5%). Socioeconomic-position-tailored interventions did not yield better outcomes compared with non-socioeconomic-position-tailored interventions for disadvantaged groups . We observed similar effect sizes in separate meta-analyses of non-socioeconomic-position-tailored interventions using trial data from participants with high socioeconomic position  and participants with low socioeconomic position , although certainty of evidence from these studies was graded as low.Of 2376 studies identified by our literature search, 348 full-text articles were retrieved and screened for eligibility. Of these, 42 studies (26 168 participants) were included in the systematic review. 30 (71%) of 42 studies were done in the USA, three (7%) were done in the UK, two (5%) each in the Netherlands and Australia, and one (2%) each in Switzerland, Sweden, Turkey, India, and China. 26 (62%) of 42 studies were trials of socioeconomic-position-tailored interventions and 16 (38%) were non-socioeconomic-position-tailored interventions. 17 (65%) of 26 socioeconomic-position-tailored interventions were in-person or telephone-delivered behavioural interventions, four (15%) were digital interventions, three (12%) involved financial incentives, and two (8%) were brief interventions. Individuals who participated in an intervention, irrespective of tailoring, were significantly more likely to quit smoking than were control participants (RR 1·56, 95% CI 1·39–1·75; IWe found evidence that individual-level interventions can assist disadvantaged smokers with quitting, but there were no large moderating effects of tailoring for disadvantaged smokers. Improvements in tailored intervention development might be necessary to achieve equity-positive smoking cessation outcomes.Cancer Research UK. In most high-income countries, tobacco smoking prevalence and the associated burden of mortality and disease7Evidence before this studyWe searched MEDLINE, PsycINFO, Embase, Cochrane Central Register, and Tobacco Addiction Register of Clinical Trials and the IC-SMOKE project database for studies published in English from database inception until Aug 18, 2019, with the following search terms: smoking cessation or smok* quit* or smok* stop* or smok* cease or smok* cessat* or smok* give up (title and abstract); systematic review or review or RCT or randomi?ed controlled trial or trial or randomi?ed or pragmatic clinical trial (title and abstract); behavio* or behavio?ral support or intervention or counsel* or brief or support or psychol* or individual* or individual-level or behavio?r therapy or cognitive therapy or target* or adapt* or tailor*) not pharma* (title and abstract); and equity or equity impact or inequalit* or poor or disparit* or SES or socio-economic or socio-economic or depriv* or disadvant* social class or occupation or employ or unemploy* or educat* or income or poverty. Tobacco control experts from the authors’ institution and others working within the UK Centre for Tobacco and Alcohol Studies were consulted about relevant submitted or in press articles. Several Cochrane reviews focused on individual-level interventions that were not tailored for low socioeconomic position, including motivational interviewing, behavioural support, and different uses of pharmacotherapy. Bauld and colleagues (2010) examined the equity effect of non-socioeconomic-position-tailored interventions. Reviews by Murray and colleagues (2009) and Bryant and colleagues (2011) focused on interventions targeted at disadvantaged smokers. These reviews suggested that, despite behavioural interventions showing promise for reducing inequalities, smoking cessation generally remains lower among disadvantaged groups. However, these reviews did not examine whether socioeconomic position tailoring moderated intervention effectiveness compared with non-socioeconomic-position-tailored approaches.Added value of this studyTo our knowledge, no previous reviews have extended examination of the overall effect of all types of individual-level interventions for smoking cessation in socioeconomically disadvantaged groups to also consider whether socioeconomic position tailoring moderates this effectiveness. We found that both socioeconomic-position-tailored and non-socioeconomic-position-tailored individual-level interventions were effective for smoking cessation in disadvantaged groups. However, there were no large moderating effects of tailoring the interventions for disadvantaged groups compared with not tailoring the interventions. This analysis is an important step forward in gathering evidence about the effectiveness of tailored approaches and encourages further research to improve the effectiveness of equity-focused smoking cessation programmes.Implications of all the available evidenceThis systematic review and meta-regression highlights the challenges in achieving improved long-term smoking cessation in disadvantaged groups through tailoring of interventions. Our results do not imply that socioeconomic-position-tailored approaches should be abandoned, but rather that to improve rates of smoking cessation among disadvantaged smokers new, multifaceted approaches are required at the individual, community, and population level, recognising the wider context of socioeconomically disadvantaged smokers. Further research should assess whether current interventions could be further adapted and improved to extend the benefits into longer-term success over and above the effectiveness of non-socioeconomic-position-tailored approaches.Regular smoking is established and maintained by a variety of molecular and behavioural factors linked to the rapid release of nicotine from cigarettes.16The terms socioeconomic position and disadvantaged were operationalised in this Article as populations facing inequalities, marginalisation, or disadvantage in terms of social class, occupation, unemployment, income, poverty, or residential neighbourhood.In theory, tailoring interventions to participant characteristics can enhance effectiveness by relating to a participant's life and needs or overcoming specific obstacles to achieve a desired change.Previous reviews have examined the equity effect of non-socioeconomic-position-tailored interventionsIf socioeconomic-position-tailored interventions are not markedly more effective than non-socioeconomic-position-tailored interventions at increasing smoking cessation among smokers with disadvantaged socioeconomic position then these approaches will require redesign. Therefore, we aimed to assess whether the effectiveness of individual-level smoking cessation interventions for disadvantaged groups was moderated by socioeconomic position tailoring.This systematic review and meta-regression followed PRISMA guidelines.This meta-analysis is based on individual participant data. Study authors were contacted if data were not available in a published report. Individual participant-level data were extracted from each study to calculate risk ratios (RRs) and 95% CIs. Studies were excluded if they were delivered at the community or population level, did not report differential effects by socioeconomic position, did not report smoking cessation outcomes from 6 months or longer after the start of the intervention, were delivered at a group level, or provided pharmacotherapy with standard behavioural support compared with behavioural support alone,protocol is available online.LK did the literature search. LK and CS independently screened all abstracts. LK screened all full-text articles and CS screened 10% of full-text articles. Inter-rater reliability at abstract screening (Cohen's κ=0·81) and full study screening (Cohen's κ=0·78) were high. Data were extracted by LK. To check reliability, 10% of data extraction was done independently by HT-B. Percentage agreement was more than 98% after comparison . ConflicDuplicate papers reporting data from the same trial were identified and the secondary papers were excluded before data extraction. We extracted data on study type and setting, participant characteristics, intervention details, and smoking cessation outcomes Diverse interventions, settings, and participants characterise the field of smoking cessation. We judged it likely that the included studies would show heterogeneity in treatment effect . As such, the assumptions of a fixed-effect meta-analysis ,In random-effects meta-analysis models (restricted maximum-likelihood method),2 test to show whether observed differences in results were compatible with chance alone. We calculated I2 statistics to examine the level of inconsistency across study findings.I2 values reflect the degree of overlap of CIs, with lower values indicating that any observed variance is spurious and higher values suggesting that there are real differences in effect size between studies. Publication bias was assessed using funnel plots. Where visual inspection indicated potential funnel plot asymmetry, we did Egger's regression test to investigate this.We explored heterogeneity by observation of forest plots and use of the χWe made the following comparisons using forest plots: individual-level interventions (tailored and not tailored to socioeconomic position) versus passive or active control or usual care; socioeconomic-position-tailored individual-level interventions versus passive or active control or usual care; and non-socioeconomic-position-tailored individual-level interventions (subgroups of low socioeconomic position and high socioeconomic position participants) versus passive or active control or usual care.vs below median), whether interventions involved a trained smoking cessation specialist (yes vs no), and use of pharmacotherapy in the intervention group (yes vs no). These covariates were included a priori as potential confounders given that programmes tailored to socioeconomic position might include more intervention sessions or components or be delivered by different professionals with varying experience. The regression coefficient estimates how the intervention effect in the socioeconomic-position-tailored subgroup differs from the reference group of non-socioeconomic-position-tailored interventions. The true effect for smoking cessation (θi) in the adjusted meta-regression is given byk is the sampling error through which the effect size of the study deviates from the true effect, and ζk indicates that the true effect size of the study is sampled from an overall distribution of effect sizes.A conventional meta-analysis attempts to combine results from studies to elucidate a single summary effect size, but diversity in populations and methods among studies often leads to statistical heterogeneity in the true effects of these studies. Meta-regression acts to extend subgroup analyses and allows, in principle, the effects of multiple factors to be investigated simultaneously. Therefore, in contrast to a meta-analysis, meta-regression aims to relate the size of effect to one or more characteristics of the studies involved. In meta-regression, a pooled effect estimate is predicted based on the values of one or more explanatory study-level variables that might influence the size of the intervention effect.Where a non-significant (p>0·05) association between socioeconomic position tailoring and intervention effectiveness was found, we used sensitivity analyses using Bayes factors to examine whether the association reflected evidence of no effect, evidence of an effect, or whether the data were insensitive to detection of an effect.36We calculated further exploratory unadjusted univariate and adjusted models to explore the extent to which important study characteristics could explain anticipated heterogeneity in the study estimates.calculator. The study is registered with PROSPERO, CRD42018103008.Analyses were done in the RStudio development environment version 1.1.463 using R version 3.5.2 and the metafor package.The funder of the study had no role in study design, data collection, data analysis, data interpretation, or writing of the report. The corresponding author had full access to all the data in the study and had final responsibility for the decision to submit for publication.Of 2376 studies identified by our literature search, 348 full-text articles were retrieved and screened for eligibility. Of these, 42 studies (26 168 participants) were included in the systematic review ; table 138–40,43,45–48,50,52–56,58–60,63,64,66–70,72–75,78,79 three (7%) were done in the UK,30 (71%) of 42 studies were done in the USA,41,46,54,55–57,59,60,65,77 Nine studies recruited only women.38,40,44,45,47,48,51,53,54,56,62,63,65,67–71,73–77 and African American participants were the majority in 12 studies.Ten studies recruited participants during hospital or clinic visits related to general health, cardiac health, dental health, or the health of a participant's child.In-person or telephone support typically included one or more sessions with a health professional who assisted in the quit attempt. These professionals included clinicians, nurses, or health educators, who either provided smoking cessation support as part of their job or worked as a smoking cessation specialist. Digital behavioural support involved interactive and tailored smoking cessation support delivered via text messages, or on a website or page accessible on a computer or other device. Financial incentive condition participants received incentives that were conditional upon them attending support sessions or health visits or contingent upon biochemically validated smoking abstinence at follow-up. Brief interventions consisted of brief advice and assistance related to smoking cessation and outlined general health risks from smoking.Overall, six (14%) of 42 included studies were classified as being at low risk of bias on all domains considered in the assessment .I2=54·5%). The certainty of evidence for this comparison was deemed to be moderate. A funnel plot suggested that there was no reporting bias for smoking cessation outcomes  indicated weak evidence that tailoring had no effect on intervention effectiveness. Repeating the calculation based on an expected effect size of 1·1 showed that the data were insensitive to detection of small effects (Bayes factor=0·81).I2=16·55%, R2adjusted=82·09%; p=0·0027) compared with the result from the primary meta-analysis , studies with some concerns in at least one domain of the Cochrane risk-of-bias tool for this result, but no high risk of bias for any domain, and pharmacotherapy had meaningful associations with effect size . An adjuI2=38·10%).We estimated a pooled effect size based on the 26 studies of socioeconomic-position-tailored interventions . SmokersWe estimated pooled effect sizes separately for participants with low socioeconomic position and participants with high economic position based on the 12 studies of non-socioeconomic-position-tailored interventions that reported outcomes . Four noI2=76·6% and 2I=82·7%, respectively; Individuals with low and high socioeconomic position who participated in a non-socioeconomic-position-tailored intervention were significantly more likely to quit smoking than were controls. However, we found evidence of high heterogeneity in the effect size between trials for both the low socioeconomic position and high socioeconomic position subgroups  rather than continued abstinence outcomes. Although there is some debate as to which is a more robust measure, a 2010 systematic reviewThis systematic review is not without limitations. Although several covariates were prespecified in the protocol, it was not possible to do the same for all other potentially important covariates and this might result in false-positive conclusions. Therefore, results indicating a reduction in heterogeneity compared with the primary meta-analysis should be viewed as exploratory. Furthermore, study characteristics included in the meta-regression might have been highly correlated, such that an observed association with one study characteristic is in fact reflective of a true association with another correlated characteristic that has not been measured. There was some evidence of clustering of study characteristics, whereby more sessions appeared to take place if a trained specialist was delivering the intervention. It is also possible that the effectiveness of behavioural support depended on the skill of the practitioner delivering it;There are potential limitations related to the operationalisation of socioeconomic position in this Article. Although 39 (93%) of 42 studies were done in high-income countries, there are often between-country differences in terms of how socioeconomic position is experienced and how this influences health behaviour.During the study screening process, it became apparent that relevant studies (n=161) might have been excluded because they did not report their outcomes by socioeconomic position, despite potentially having a socioeconomically diverse sample of participants. Given the persistent inequalities in smoking rates worldwide, it is becoming ever more important that smoking cessation trials, where possible, collect and report outcomes by socioeconomic position. Studies are typically not powered for robust subgroup analyses by socioeconomic position, but if outcomes are reported in this way then they can be cumulatively included in pooled effect size estimates in future reviews. The certainty of evidence of studies included in this review was rated as moderate for the primary analysis and low for the secondary analyses. As such, it remains possible that the true effects are different to what was estimated.Despite these limitations, this study has several strengths. To our knowledge, no previous reviews have examined whether socioeconomic position tailoring moderates the effectiveness of individual-level behavioural smoking cessation interventions at 6 months or later in socioeconomically disadvantaged groups. Inclusion of 42 studies in our systematic review made it possible to do a meta-regression analysis, which is a useful tool to extend the analysis and relate the size of treatment effect in clinically and methodologically diverse studies to relevant study characteristics. Considering the growing number of interventions that involve some form of tailoring for disadvantaged groups, this analysis is an important step towards gathering evidence about their effectiveness and might also encourage further equity-focused research that will improve the effectiveness of smoking cessation programmes. Future research in this area should also consider assessing what the most effective components of socioeconomic-position-tailored interventions are by using an appropriate theory-informed taxonomy.84This systematic review and meta-regression highlights that although both socioeconomic-position-tailored and non-socioeconomic-position-tailored individual-level interventions for smoking cessation in socioeconomically disadvantaged groups are effective for smoking cessation, based on the evidence available for this review, there is currently no evidence for large moderating effects of tailoring for disadvantaged smokers.


Feline infectious peritonitis (FIP) is a coronavirus-induced disease of cats, in which the immune system is known to play a crucial, but complex, role in the pathogenesis. This role is still incompletely understood, with involvement of both host and viral factors. To evaluate differential gene expression and pathway involvement in feline coronavirus (FCoV) infection and FIP, we applied next-generation RNA-sequencing of the mesenteric lymph nodes from cats with naturally-acquired FIP, as well as those with systemic FCoV infection without FIP, and those with neither. Viral infection was associated with upregulation of viral defenses regardless of the disease state, but to a greater degree in FIP. FIP was associated with higher pro-inflammatory pathway enrichment, whilst non-FIP FCoV-positive cats showed lower enrichment of humoral immunity pathways, below that of uninfected cats in the case of immunoglobulin production pathways. This host response is presumed to be protective. In FIP, downregulation of T cell-related processes was observed, which did not occur in non-FIP FCoV-positive cats. These results emphasize the importance of the host’s immune balance in determining the outcome of the FCoV infection. Orthocoronavirinae, it belongs to a different subfamily than known nidoviruses of reptiles and mammals [Feline infectious peritonitis (FIP) is a fatal immune-mediated disease of domestic and wild felids caused by feline coronavirus (FCoV), a highly prevalent virus with a worldwide distribution which infects domestic and wild felids. FCoV is an alphacoronavirus, similar to several respiratory and enteric pathogens in other species, such as canine coronavirus (CoV), porcine transmissible gastroenteritis CoV, and the ‘common cold’ CoVs of humans. It differs at the genus level from the betacoronaviruses that include the severe respiratory disease-causing CoVs of humans , and the gammacoronaviruses of birds. As a member of the subfamily  mammals ; this ma mammals . In ~5%  mammals . FIP is  mammals ,8,9.Numerous studies have tried to pinpoint a key viral mutation leading to the disease of FIP. However, this has so far proven inconclusive ,11. CertSusceptibility to disease has been shown to be heritable to some extent, suggesting a genetic component ,20,21. AThe hallmark pathological processes of FIP, monocyte-mediated granulomatous phlebitis and systemic endothelial activation, the changes in the lymphatic tissue (such as lymphocyte depletion) together with clinical signs such as fever, all indicate an excessive and inappropriate immune response ,28,29. TUtilizing a rare biobank from cats with FIP, as well as FCoV-infected and uninfected cats without FIP, we previously evaluated mesenteric lymph nodes (MLN) for expression of selected immune mediators and found a possible intermediate stage of activation, represented by upregulation of some mediators, in FCoV-infected cats without FIP . Natural4 at the closest values).Reverse-transcriptase-quantitative PCR (RT-qPCR) results for FCoV in the MLN of these cases have been previously published ,13,14 anThe sequencing of the RNA libraries generated on average 25.3 Mio reads . The mappability of these reads was high (89.7% on average) and remarkably consistent, with a narrow range of 87.4 to 89.8%. Between 59 and 72% of the mapped reads were assigned to coding regions and contributed to the estimation of the gene expression levels. Together with the summary of the quantification and the pair-wise comparisons, this is shown in Based on the expression profiles, a clear separation between FIP cats and non-FIP cats was obtained by means of multidimensional scaling (MDS) . To a lep-value below 0.05 and a log2ratio above (or below) 0.5. These sets of up or down regulated genes were used as inputs for the GO databases.In order to perform an ensemble functional analysis of the differences in the transcriptional patterns between the groups, we selected from the pair-wise comparisons genes with a Functional analysis of the genes significantly upregulated in FIP reveals 83 enriched gene ontology (GO) terms associated to a biological process (BP) . They alWhen individual genes are looked at, and ranked by fold-change between groups, the top 20 upregulated in FIP includes IRF7, IFIT2&3, and ISG15  as well as C5, CCL8, and CXCL10 .A single BP GO term is higher in non-FIP cats: Peptidyl-tyrosine autophosphorylation; however, the CC term GO:0042101 (T cell receptor complex) is also higher in these cats.The functional analysis of differentially expressed genes between these groups reveals that the significantly enriched GO terms are again almost exclusively driven by genes upregulated in the MLN of FIP cats, with three exceptions . The mosComparing the MLN data from the subgroups of non-FIP cats  resulted in far fewer DE genes and hence fewer significantly enriched GO terms  were themselves compared to identify the extent of differences in the MLN response when infection outcome  was considered. A Venn diagram  shows thA subset of GO biological processes deemed of most interest was further analyzed in order to compare ensemble gene expression within categories between groups see . ConsideIn most cases the gene expression levels were higher in the MLN from cats with FIP (G2) than in both non-FIP subgroups . GO:006910 (phagocytosis recognition) additionally showed significantly higher levels in FCoV-negative than in FCoV-positive non-FIP cats; within GO:0032481 (positive regulation of type I interferon production) gene levels between the two non-FIP groups were found to be slightly higher but not significantly so in FCoV-positive cats. The two T cell-related categories  both had gene levels in MLN from FIP cats significantly lower than in the non-FIP subgroups, which were themselves not significantly different from each other. Genes of GO:0032755 (positive regulation of interleukin-6 production) were higher in MLN from FIP cats than in either non-FIP subgroup, which were again similar between themselves. Immunoglobulin production category genes (GO:0002377), though higher in the MLN from FIP cats, were lower in FCoV-positive than in FCoV-negative non-FIP cats.The results of the present study for the first time provide detailed information on how the immune system responds to FCoV infection in natural cases, both in animals succumbing to FIP and those showing no clinical or pathological signs attributable to FIP. In focusing on the MLN, we chose to study an organ that is consistently involved in FCoV infection , as wellThousands of genes were found to be dysregulated in the MLN of cats with FIP, pointing to numerous immune and inflammatory pathways, most of which were predictable or consistent with known features of the disease. Viral defense and negative regulation of viral replication were found to be enriched in FCoV infection regardless of FIP status, but at significantly higher levels in cats with FIP, demonstrating that the host attempts, albeit unsuccessfully in cats with FIP, to respond appropriately. Similarly, processes related to the prototypic anti-viral cytokines, type I and II IFN were also enriched in FCoV infection regardless of FIP status, whilst IFN pathway genes were heavily overrepresented amongst the top DE genes found to be higher . HoweverWe also identified that bacterial response pathways were upregulated in the MLN of cats with FIP. There are two likely explanations for this observation. One is an overlap in function between pattern recognition receptors (PRR) responsible for detecting pathogens, e.g., TLRs 2, 4, and 8 were all upregulated in the MLN of cats with FIP and yet not in the MLN from FCoV-positive cats without FIP. This fits with previously reported findings . TLR2 is+ regulatory T (Treg) cells [reg cells are crucial for maintaining immune homeostasis and can also suppress other leukocytes, a function which would be expected to be beneficial in cats with an exuberant immune response [reg cells, and hence their immune regulatory effects, has also been shown in the peripheral blood of cats with FIP [There were few processes found to be significantly downregulated in the MLN of cats with FIP, but those that were have major immunological implications. These included T cell receptor signaling pathway, and IL-4 production. When genes from the categories T cell differentiation and T cell mediated immunity were analysed separately, T cell features were relatively lower in FIP cats than either group of cats without FIP. Consistent with this, anti-viral T cell responses (as determined by proliferation assays and T cell subsets) were found to be higher in cats resistant to experimental FIP in a recent study . This sug) cells . Treg ceresponse . A reducwith FIP . IL-4 iswith FIP ,41. It hwith FIP .Results in FCoV-positive cats without FIP show that many of the same processes as in cats with FIP are significantly upregulated, as compared to FCoV-negative cats without FIP, including viral defense, leukocyte chemotaxis, IL-1 response, and IFN-response pathways. However, a closer look at the individual gene levels of these processes shows that they are higher in cats with FIP than in both groups of cats without FIP. This suggests that quantitative differences may be key. Indeed, in both FCoV-positive groups  there was upregulation of anti-viral IFN-related genes but the inflammatory triad of IL-1, IL-6, and TNF, as well as CXCL10, a highly inflammatory chemokine, were only elevated in the MLN of cats with FIP. Therefore, these are not likely to have any positive contribution to viral defense and rather act to induce the observed pathological processes and clinical signs. Despite the upregulation of type I IFN pathways, IFN-α and-β themselves were not significantly upregulated in the MLN of cats with FIP, suggesting that further investigation into defects in this pathway may be required. Transcription factors for IFN pathways (including IFN-regulatory factors (IRF7 and 9), as well as IFN-stimulated genes  and IFN-induced proteins (IFIT2 and 3) were all elevated in both FCoV-positive groups. Members of the large (>70 member) tripartite motif-containing (TRIM) family, some of which are involved in innate antiviral immunity , were upCrucially, there is less upregulation of humoral immunity  in FCoV-positive cats without FIP in contrast to cats with FIP, further supporting the early theories that this branch of the immune system is detrimental ,45. In cPrevious studies have found apoptosis to be a prominent feature in FIP, both morphologically in the lymphoid tissue and at the signaling pathway level in macrophages ,41,42. IThe study population consists of animals presented to either the university small animal hospitals or local practices in Zurich, Switzerland, and Bristol, UK. All animals, with or without FIP, were euthanized for purely clinical reasons. Cases are a subset of those previously published with the exception of sample 2.9. University of Bristol FIP Biobank samples were collected at post-mortem examination with full informed consent from owners that samples could be used for research purposes. The collection, storage and use of samples used in this project were approved under ethical review by the University of Bristol Animal Welfare and Ethical Review Board  ,13,14. C® (Qiagen) with on column genomic DNA (gDNA) removal was used following the manufacturer’s instructions. Briefly, approximately 30 mg of tissue was thawed on ice and RNA extraction buffer was added. Following disruption and homogenization using a tissue homogenizer , RNA was extracted and eluted. Remaining gDNA was removed by an additional DNase step . Complementary (cDNA) synthesis was performed with the High-Capacity cDNA Reverse Transcription Kit (Thermo Fisher Scientific) according to the manufacturer’s protocol. Before conducting RT-qPCR, the RNA levels of the samples were adjusted to 400 ng/μL, by measuring the concentration with a NanoDrop 2000® (Thermo Fisher Scientific) and further diluted 1 in 20. TaqMan RT-qPCR was performed as previously described using a Biosystems 7500 Fast PCR System® (Thermo Fisher Scientific) and published protocols [For RNA extraction, the RNeasy Plus Minikitrotocols .® (1.0) Fluorometer  and a Bioanalyzer 2100 were used to determine the RNA quality; only samples with a 260/280 nm ratio between 1.8–2.1 and a 28S/18S ratio within 1.5–2 were processed. Following the manufacturer’s protocol, the stranded (antisense) TruSeq RNA Sample Prep Kit v2  was used. After polyA enrichment of total RNA, the samples were reverse-transcribed into cDNA. This was followed by fragmentation, end-repair, and polyadenylation, before ligation of the TruSeq adapters (containing an index). This allows multiplexing fragments and selective enrichment by PCR. For validating the quality and quantity of the libraries, the Qubit® (1.0) Fluorometer and the Caliper GX LabChip® GX  were used. The product is a smear with an average fragment size of approximately 260 bp. The libraries were normalized to 10 nM in Tris-Cl 10 mM, pH 8.5 with 0.1% Tween 20.A QubitAfter library quantification, libraries were prepared for loading according to the NovaSeq XP workflow with the NovaSeq6000 SP Reagent Kit . Cluster generation and sequencing were performed on a NovaSeq6000 System with a run configuration of single end 100 bp.Felis catus  with STAR v 2.7.3 [Reads were quality-checked with FastQC . Sequencing adapters were removed with Trimmomatic  and alig v 2.7.3  using thGenomicRanges [Distribution of the reads across genomic isoform expression was quantified using the R package icRanges . MinimumedgeR [present and a test was performed. Of those, genes with a p < 0.01 and a log2 fold-ratio > 0.5 (<−0.5) were marked as differentially upregulated (respectively downregulated).Differentially expressed (DE) genes were identified using the R package edgeR , using aenricher, from the package clusterProfiler [Enrichment of GO terms was performed using the hypergeometric test as implemented in the R function Profiler .plotMDS from the package limma [pheatmap [The multi-dimensional scaling plot was generated with the R function ge limma  and the pheatmap .adonis2 from the package Vegan [The potential effect of additional covariates has been tested using PERMANOVA (with 999 permutations) as implemented in the R function ge Vegan .All R packages used were from Bioconductor Version 3.10.For all pairwise comparisons, the enrichment plots show the top 10 most enriched biological processes (GO BP) based on significantly up or downregulated genes. Tables with the full lists of significantly enriched GO terms are provided in the In conclusion, our findings highlight the difference in host response to FCoV in the context of systemic infection both in cats with and without FIP. They demonstrate the extent of immune system activation in FIP as well as the skew towards a pro-inflammatory cytokine response and further demonstrate the importance of a strong T cell-mediated immunity in mounting an effective response to FCoV infection. Comparisons of host response to infection with different clinical outcomes  also support the initial hypothesis that an excessive pro-inflammatory response is responsible for FIP.


Acinetobacter baumannii isolates due to the acquisition of blaOXA-23 in a regional hospital located in Brazilian Amazon Region.  In this study, we report a clonal dissemination of carbapenem resistant  The isolates were identified by MALDI-TOF and the carbapenemase-encoding genes were detected by multiplex-PCR. The genetic similarity was investigated by pulsed-field gel electrophoresis (PFGE). blaOXA-23. PFGE analysis revealed that these isolates belong to a single clone.  Only 10 (55.6%) isolates harbored the gene  This dissemination strategy indicates the need for surveillance, adoption of control procedures defined in guidelines, and the careful administration of antimicrobials should be reinforced. Acinetobacter baumannii is an opportunistic pathogen with several virulence factors associated with several outbreaks worldwide, especially among intensive care unit (ICU) patients with severe underlying diseases,,A. baumannii, led the World Health Organization (WHO) to gather global leaders at the United Nations (UN) General Assembly meeting in 2016 to commit them to fight against antimicrobial resistanceThe worldwide emergence and dissemination of MDR bacteria, such as A. baumannii, genes encoding resistance to multiple broad-spectrum antimicrobials are commonly detected, including carbapenem resistance genesA. baumannii isolates increased by approximately 60% compared to resistance between the periods of 1977-1999 (12.6%) and 2008-2010 (71.4%)In infections caused by A. baumannii (MDR-AB) is a risk factor for the dissemination of this pathogen among patients and for the subsequent development of infectionsIn addition, it should be noted that environmental and patient colonization by MDR Hence, MDR-AB isolates are associated with high mortality rates, costs, and dissemination among hospitalized patients. Studying MDR-AB dissemination is important for implementing effective infection control/colonization measures, breaking the epidemiological chain of transmission of this microorganism, mitigating rates of bacterial resistance, reducing morbidity and mortality, and improving the quality of healthcare.Considering these factors and the lack of research on this topic in the Brazilian Amazon region, the current study aimed to determine the phenotypic and genotypic characteristics of MDR-AB isolates, to detect the presence of carbapenemases, and to demonstrate the genetic similarity among carbapenemase-producing isolates at a tertiary referral hospital in the Amazon.A cross-sectional descriptive study was performed between September 2017 and February 2018 at a regional public hospital in the southeast region of Pará state (Brazil). This is a tertiary referral hospital with different medical specialties and with pediatric and adult intensive care units, particularly for nephrology, with kidney transplantation and renal replacement therapy services. The hospital treats an estimated population of 541,000 inhabitants and is located in the Amazon biome. A. baumannii resistant to imipenem and meropenem were included in the study. Only one isolate per patient was analyzed. The isolates were obtained from clinical samples including tracheal secretions, blood cultures, postoperative wound swabs, catheter tips, and inguinal swabs. A total of 18 bacterial isolates were recovered from patients undergoing treatment at the regional hospital and diagnosed with a hospital-acquired infection (HAI) or with colonization caused by Antimicrobial susceptibility was performed using the disk diffusion test and the results were interpreted according to the Brazilian Committee on Antimicrobial Susceptibility Testing (BrCAST) guidelines® strip (bioMérieux) for isolates resistant to imipenem and also using the disk diffusion test for meropenem. MIC of polymyxin B  was determined using the broth microdilution susceptibility test . The concentrations tested ranged from 0.125 to 64 µg/mL and the bacteria were considered resistant when the MIC values were ≥ 4 µg/mLThe minimum inhibitory concentration (MIC) of carbapenems was defined using the ETEST® 96 plus  and interpreted according to the thresholds set by the Food and Drug Administration (FDA) . The standard strain E. coli ATCC 35218 was used as a control. The kits Carbapenembac™ and Carbapenembac-Metalo™ (PROBAC®) were used for phenotypic detection of carbapenemases and confirmation of positive isolates, respectively. MIC for tigecyclines was determined using the MicroScan AlkAway®- QIAGEN), containing 25 ng of DNA and 0.5 µL of each primerblaVIM, blaSPM, blaGIM, blaIMP, blaSIM, blaNDM, blaKPC, blaOXA-23, blaOXA-24, blaOXA-51, blaOXA-58, and blaOXA-143The identification of all positive isolates in the phenotypic test for carbapenemases (n = 10) was confirmed by MALDI-TOF. For genotyping the isolates, chromosomal DNA was extracted from 3-5 colonies after boiling for 15 min and centrifuged at 12,000 rpm at room temperature. PCR was performed in a final volume of 25 µL of master mix (TopTaqA. baumannii isolates was determined by pulsed-field gel electrophoresis (PFGE), using the molecular weight marker Lambda PFGE Ladder® (GelSyringe ™). Isolates were sent to laboratory of Alerta at the Federal University of São Paulo (UFSP) for molecular typing. For this purpose, bacterial suspensions were digested with the restriction enzyme ApaI  and the DNA fragments were separated by 1% agarose gel electrophoresis  in 0.5X TBE buffer . Electrophoresis was performed using the CHEF-DR II system  at 14 ºC, using 200 volts (6 V/cm) electric current with an initial switch time of 5 s and a final switch time of 35 s for 19 hrs. The gel was stained with UniSafe Dye®  and photographed under ultraviolet light. PFGE-stained photos and DNA fragments were examined using BioNumerics software version 5.0 Genetic similarity between Samples were obtained from ICU patients (50%) and ward patients (50%). The isolates were collected from tracheal secretions , blood cultures , postoperative wound swabs , catheter tips , and inguinal swabs . Among the 18 isolates included in the study, the majority of patients were male , whose ages ranged from 18 to 86 years. The isolates were resistant to almost all antimicrobials tested, and remained susceptible to tigecycline and polymyxin B. As shown in Table1, one isolate was not tested for polymyxin B and tigecycline due to loss of cell viability during the susceptibility test .blaOXA-23 and blaOXA-51 gene. PFGE analysis showed that all isolates which tested positive for carbapenemase belonged to a single clone  which were carbapenemase-producing. Three of them (16.7%) were positive for metalo-β-lactamases and seven (83.3%) for serine carbapenemases. However, genotypic testing did not confirm these findings . In all le clone .--MDR-AB is more frequently isolated from colonized and/or HAI patients, especially in hospital environmentsIn this study, most of the isolates were susceptible to tigecycline. Although this antimicrobial has been licensed for treating complicated intra-abdominal and skin infections and community-based bacterial pneumonia, it has been widely used off-label to treat many other infections, including those caused by MDR-ABA. baumannii isolates in this study were collected from colonized ICU patients. It should be noted that either patient colonization or infection with MDR-AB are important sources of dissemination of resistant strains between hospitalsMost blaOXA-23 and blaOXA-51 were detected in all carbapenemase-producing isolates, similar to what has been described in a previous study, which reported a carbapenem resistance rate of 90%A. baumannii isolates that are endemic in several Brazilian statesThe genes A. baumannii, but less effective than other enzymatic mechanismsOXA-23 have high MICs for imipenem and meropenem, but those expressing only blaOXA-51 have a lower MIC due to the reduced hydrolytic activity of OXA-51 for carbapenens,The oxacillinase (OXA)-type carbapenemase expression is the most common mechanism of resistance in blaOXA-23-encoding A. baumannii strains, matching previous studies on ICU isolates which reported 91.8% and 100% of genetic identity,A. baumannii isolates related to patient infection or colonization, whose source of dissemination could have been health staff, equipment, or contaminated fomites,The PFGE pattern demonstrated a high genetic similarity and dissemination of ,Preventing the clonal dissemination of microorganisms and improvement in the healthcare quality requires strategic prevention efforts and well-known infection control practices. For example, applying hand hygiene and equipment cleaning rules and creating staff awareness are particularly important to prevent the spread of an infectionA. baumannii isolates studied, and that all isolates had MDR bacteria. The genes blaOXA-23 and blaOXA-51 were detected in all carbapenemase-producing isolates, and all these isolates belonged to the same clone. Of all the antimicrobials tested, polymyxin B and tigecycline were the most effective antimicrobials for MDR-AB. The limitations identified in our study were the relatively small number of isolates from patients, who had several comorbidities and some of the patients were hospitalized multiple times; and the limited quality of the clinical data, which was insufficient to meet the study aims. The results of this study demonstrate that carbapenem resistance was common to all A. baumannii and rational administration of antimicrobials should be reinforced.Bacterial resistance is an emerging problem requiring the utmost attention and effort towards its mitigation. Our study highlights the need for screening colonized or infected patients and for providing frequent training to healthcare professionals in both ICUs and clinics. Moreover, surveillance for imipenem- and meropenem-resistant