Abstract T MP77: Incidence of Neurofluctuation in Patients with Subcortical Ischemic Stroke - Results from a Prospective Cohort Study
Farhaan VahidyWilliam J. HicksIndrani AcostaHen HalleviHui PengRenganayaki PandurenganNicole R. GonzalesAndrew D. BarretoSheryl Martin‐SchildTzu‐Ching WuMohammad H. RahbarArvind BambhroliyaJames C. GrottaSean I. Savitz
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Subcortical strokes (SCS) are associated with stroke progression. The primary objective was to assess the incidences of neurological deterioration and its spontaneous reversal in SCS patients. We also estimated the risk of poor outcome associated with deterioration. Methodology: A prospective cohort of SCS patients was enrolled based on clinical exam and imaging studies, and was followed through the course of hospitalization. NIH stroke scale (NIHSS) assessments were performed daily and whenever symptoms worsened (increase of ≥ 1 in motor scale). Modified Rankin Scale (mRS) score at discharge was used to assess outcomes. Patients received standard of care stroke therapy including IV t-PA as per AHA guidelines, and episodes of deterioration were managed conservatively. The Anderson Gill Cox proportional hazard model was used to analyze factors associated with time to deterioration and spontaneous reversal. Log binomial regression was performed for estimation of risk of poor outcome. Results: The cumulative incidence of deterioration was 43.3% and of those who deteriorated, spontaneous reversal occurred in 37.8%. The median time to deterioration was 21 hours from stroke onset. Use of IV t-PA was significantly associated with both deterioration (HR: 2.25, 95% CI: 1.13 - 4.49) (Figure 1a) and spontaneous reversal (HR: 4.36, 95% CI: 1.36 - 14.01) (Figure 1b). Deteriorated patients had a higher risk of poor discharge outcome (mRS ≥ 3) (RR: 1.80, 95% CI: 1.71 - 1.93), after adjusting for age, initial stroke severity, and use of IV t-PA. Conclusion: The results from this study establish a natural history of neurofluctuation in patients with SCS. Our data suggests that deterioration occurs early in these patients, and only about a third of deteriorated patients reverse spontaneously. Furthermore, deterioration is associated with poor short term functional outcome. There is a need to develop and test novel therapies to treat neurological deterioration in SCS.Keywords:
Stroke
Cumulative incidence
A perceived weakness of the modified Rankin Scale is potential for interobserver variability. We undertook a systematic review of modified Rankin Scale reliability studies.Two researchers independently reviewed the literature. Crossdisciplinary electronic databases were interrogated using the following key words: Stroke*; Cerebrovasc*; Modified Rankin*; Rankin Scale*; Oxford Handicap*; Observer variation*. Data were extracted according to prespecified criteria with decisions on inclusion by consensus.From 3461 titles, 10 studies (587 patients) were included. Reliability of modified Rankin Scale varied from weighted kappa=0.95 to kappa=0.25. Overall reliability of mRS was kappa=0.46; weighted kappa=0.90 (traditional modified Rankin Scale) and kappa=0.62; weighted kappa=0.87 (structured interview).There remains uncertainty regarding modified Rankin Scale reliability. Interobserver studies closest in design to large-scale clinical trials demonstrate potentially significant interobserver variability.
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The utility-weighted modified Rankin Scale, representing patient perspectives of quality of life, is a newly proposed measure to improve the interpretability of the modified Rankin Scale. Despite obvious advantages, such weighting imperfectly reflects the multidimensional patterns of post-stroke burden.To investigate multidimensional patterns of post-stroke burden formed by individual domains of Assessment of Quality of Life and Barthel Index for each modified Rankin Scale category.In the A Very Early Rehabilitation Trial (n = 2104), modified Rankin Scale scores and modified Rankin Scale-stratified Barthel Index scores of Self-care and Mobility, and Assessment of Quality of Life scores of Independent Living, Senses, Mental Health and Relationships were collected at three months. The multivariate relationship between individual Assessment of Quality of Life and Barthel Index domains, and modified Rankin Scale was investigated using random effects linear regression models with respective interaction terms.Of 2104 patients, simultaneously collected Assessment of Quality of Life, Barthel Index and modified Rankin Scale scores at three months were available in 1870 patients. While individual Assessment of Quality of Life and Barthel Index domain scores decreased significantly as modified Rankin Scale increased (p < 0.0001), the patterns of decrease differed by domains (p < 0.0001). Patients with modified Rankin Scale 0-1 had the largest post-stroke burden in the Mental Health and Relationship domains, while patients with modified Rankin Scale >3 showed the greatest burden in Independent Living, Mobility and Self-care domains.Across the modified Rankin Scale, individual domains are varyingly impacted demonstrating unique patterns of post-stroke burden, which facilitates appropriate assessment, articulation and interpretation of the modified Rankin Scale and utility-weighted modified Rankin Scale.
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Background Resource use in the acute and subacute phases after stroke depends on the degree of disability. Aims To determine if direct costs after stroke also vary by level of disability as measured using the modified Rankin scale at the chronic stage after stroke. Methods In a multicentre study, we collected acute and chronic in- and outpatient resource use in survivors of ischemic stroke stratified by levels of disability according to the modified Rankin Scale. Statistical inference on costs at each level of the modified Rankin Scale was estimated using a general linear model for the first three months, the first year, and any subsequent year after ischemic stroke. Results A total of 569 survivors of ischemic stroke with a mean age of 71.7 years were enrolled (41% female) from 10 academic and nonacademic centers. Costs varied substantially over time and with each modified Rankin Scale level. The total average costs in the first year were estimated $33,147 per patient, ranging from $9,114 for modified Rankin Scale 0 to $83,236 for modified Rankin Scale 5. In the second year, medical costs were on average $14,039, varying from $2,921 to $39,723 for patients with modified Rankin Scale 0–5. The level of disability based on the modified Rankin Scale was a major determinant of resource use, irrespective of age, gender, atrial fibrillation, and vascular risk factors. Conclusion Long-term resource use after stroke is high and is mainly driven by degree of disability as measured by the modified Rankin scale.
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Background Although most people with relapsing onset multiple sclerosis (R-MS) eventually transition to secondary progressive multiple sclerosis (SPMS), little is known about disability progression in SPMS. Methods All R-MS patients in the Cardiff MS registry were included. Cox proportional hazards regression was used to examine a) hazard of converting to SPMS and b) hazard of attaining EDSS 6.0 and 8.0 in SPMS. Results 1611 R-MS patients were included. Older age at MS onset (hazard ratio [HR] 1.02, 95%CI 1.01–1.03), male sex (HR 1.71, 95%CI 1.41–2.08), and residual disability after onset (HR 1.38, 95%CI 1.11–1.71) were asso- ciated with increased hazard of SPMS. Male sex (EDSS 6.0 HR 1.41 [1.04–1.90], EDSS 8.0 HR 1.75 [1.14–2.69]) and higher EDSS at SPMS onset (EDSS 6.0 HR 1.31 [1.17–1.46]; EDSS 8.0 HR 1.38 [1.19–1.61]) were associated with increased hazard of reaching disability milestones, while older age at SPMS was associated with a lower hazard of progression (EDSS 6.0 HR 0.94 [0.92–0.96]; EDSS 8.0: HR 0.92 [0.90–0.95]). Conclusions Different factors are associated with hazard of SPMS compared to hazard of disability progres- sion after SPMS onset. These data may be used to plan services, and provide a baseline for comparison for future interventional studies and has relevance for new treatments for SPMS RobertsonNP@cardiff.ac.uk
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Abstract Background The role of allogeneic hematopoietic stem cell transplantation (allo‐HSCT) in patients <3 years of age remains controversial. Data on haploidentical donor (HID) transplants in this age group is limited. Patients and Methods We retrospectively analyzed the prognosis of 97 patients with acute leukemia aged <3 years who underwent HID transplantation at our institute. Results With a median follow‐up of 45 months, the 3‐year disease‐free survival (DFS), overall survival (OS), and 3‐year cumulative incidence rate of treatment‐related mortality were 69.3% (95% confidence interval (CI): 59.9%–78.7%), 74.2% (95% CI: 65.2%–83.2%), and 3.6% (95% CI: 0.9%–9.7%) in all 97 patients, respectively. The 3‐year DFS and OS rate in patients diagnosed <1 year and patients diagnosed ≥1 year were comparable: 77.8% (95% CI: 62.2%–93.4%) versus 66.3% (95% CI: 55.0%–77.6%, p = .253) and 82.5% (95% CI: 66.3–98.7%) versus 72.8% (95% CI: 61.9%–83.7%, p = .153), respectively. At the last follow‐up, 23 patients had died, and 20 had died of relapse. Multivariate analysis revealed that positive pre‐HSCT flow cytometric minimal residual disease (hazard ratio 5.605, p = .000) and AML‐M7 expression (hazard ratio 2.906, p = .014) were independent adverse prognostic variables for relapse. Conclusions HID transplantation is potent and safe for infants and young patients with acute leukemia. Relapse is the primary cause of treatment failure.
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The hazard ratio and median survival time are the routine indicators in survival analysis. We briefly introduced the relationship between hazard ratio and median survival time and the role of proportional hazard assumption. We compared 110 pairs of hazard ratio and median survival time ratio in 58 articles and demonstrated the reasons for the difference by examples. The results showed that the hazard ratio estimated by the Cox regression model is unreasonable and not equivalent to median survival time ratio when the proportional hazard assumption is not met. Therefore, before performing the Cox regression model, the proportional hazard assumption should be tested first. If proportional hazard assumption is met, Cox regression model can be used; if proportional hazard assumption is not met, restricted mean survival times is suggested.风险比（hazard ratio，HR）和中位生存时间是生存分析时的常规分析和报告指标。本文简要介绍了HR和中位生存时间的关系以及比例风险假定在这两者之间的作用，分析了检索出的58篇文献中的110对风险比和中位生存时间比的差异，并通过实例阐明了产生这种差异的原因。结果表明，在不满足比例风险假定时，Cox回归模型计算得到的风险比是不合理的，且与中位生存时间之比不等价。因此，在使用Cox回归模型前，应先进行比例风险假定的检验，只有符合比例风险假定时才能使用该模型；当不符合比例风险假定时，建议使用限制性平均生存时间。.
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The modified Rankin Scale (mRS) at 90 days after stroke onset has become the preferred outcome measure in acute stroke trials, including recent trials of interventional therapies. Reporting the range of modified Rankin Scale scores as a paired horizontal stacked bar graph (colloquially known as "Grotta bars") has become the conventional method of visualizing modified Rankin Scale results. Grotta bars readily illustrate the levels of the ordinal modified Rankin Scale in which benefit may have occurred. However, complementing the available graphical information by including additional features to convey statistical significance may be advantageous. We propose a modification of the horizontal stacked bar graph with illustrative examples. In this suggested modification, the line joining the segments of the bar graph (e.g. modified Rankin Scale 1-2 in treatment arm to modified Rankin Scale 1-2 in control arm) is given a color and thickness based on the p-value of the result at that level (in this example, the p-value of modified Rankin Scale 0-1 vs. 2-6)-a thick green line for p-values <0.01, thin green for p-values of 0.01 to <0.05, gray for 0.05 to <0.10, thin red for 0.10 to <0.90, and thick red for p-values ≥0.90 or outcome favoring the control group. Illustrative examples from four recent trials (ESCAPE, SWIFT-PRIME, IST-3, ASTER) are shown to demonstrate the range of significant and non-significant effects that can be captured using this proposed method. By formalizing a display of outcomes which includes statistical tests of all possible dichotomizations of the Rankin scale, this approach also encourages pre-specification of such hypotheses. Prespecifying tests of all six dichotomizations of the Rankin scale provides all possible statistical information in an a priori fashion. Since the result of our proposed approach is six distinct dichotomized tests in addition to a primary test, e.g. of the ordinal Rankin shift, it may be prudent to account for multiplicity in testing by using dichotomized p-values only after adjustment, such as by the Bonferroni or Hochberg-Holm methods. Whether p-values are nominal or adjusted may be left to the discretion of the presenter as long as the presence or absence is clearly stated in the statistical methods. Our proposed modification results in a visually intuitive summary of both the size of the effect-represented by the matched bars and their connecting segments-as well as its statistical relevance.
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Background Cognitive or communication issues may preclude direct modified Rankin Scale interview, necessitating interview with a suitable surrogate. The clinimetric properties of this proxy modified Rankin Scale assessment have not been described. Aims To describe reliability of proxy-derived modified Rankin Scale and compare with traditional direct patient interview. Methods Researchers assessed consenting stroke inpatients and their proxies using a nonstructured modified Rankin Scale approach. Paired interviewers (trained in modified Rankin Scale) performed independent and blinded modified Rankin Scale assessment of patients and appropriate proxies. Interobserver variability and agreement between patient and proxy modified Rankin Scale were described using kappa statistics ( k, 95% confidence interval) and percentage agreement. Results Ninety-seven stroke survivors were assessed. Proxies were family members ( n = 29), nurses ( n = 50), or physiotherapists ( n = 25). Median modified Rankin Scale from both patient and proxies was 3 [interquartile range (IQR): 2–4]. Reliability for patient modified Rankin Scale interview was weighted kappa = 0.70 (95% confidence interval: 0.30–1.00). Reliability for proxy modified Rankin Scale weighted kappa = 0.62 (95% confidence interval: 0.34–0.90). Subgroup analysis of various proxy information sources were as follows: family weighted kappa = 0.61; nurse weighted kappa = 0.58; therapist weighted kappa = 0.58. There was disagreement between patient-derived modified Rankin Scale and corresponding proxy modified Rankin Scale weighted kappa = 0.64 (95% CI: 0.42–0.86). Conclusions There is potential for substantial interobserver variability in proxy modified Rankin Scale and validity of certain proxy assessments is questionable. Direct modified Rankin Scale interview is preferred.
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Clinical neurology
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