Abstract
Objective
This study aims to evaluate the prognostic value of the ratio of intracerebral hematoma volume (IHV) to cerebrospinal fluid volume (CSFV) in patients with basal ganglia hemorrhage undergoing non-surgical treatment. By providing a quantitative reference for clinical decision-making, this research seeks to facilitate the early identification of patients with poor prognoses and to support the selection of optimal treatment strategies tailored to individual needs.
Methods
This retrospective study included patients with spontaneous basal ganglia hemorrhage who received non-surgical treatment at our hospital between January 2017 and October 2020. IHV, CSFV, and the maximum transverse diameter of the third ventricle (MTDTV) were measured, and baseline characteristics were collected. Prognosis was assessed at three months using the modified Rankin Scale (mRS), with scores ≤2 indicating good outcomes and scores ≥3 indicating poor outcomes. Correlations between clinical variables and prognosis were analyzed.
Results
A total of 125 patients were included in this study, among whom 74 (59.2%) experienced poor outcomes. Univariate analysis revealed significant differences in NIHSS score, GCS score, IHV, systolic blood pressure (SBP) and diastolic blood pressure (DBP) at admission, CSFV, IHV/CSFV, and MTDTV (P<0.05). Logistic regression analysis identified IHV/CSFV (OR=1.074, P=0.041) and SBP at admission (OR=1.052, P=0.005) as independent predictors of patient outcomes.IHV/CSFV demonstrated good discriminatory performance and may reflect the balance between hematoma burden and intracranial CSF reserve, potentially complementing established predictors. The area under the ROC curve for IHV/CSFV was 0.810 (95% CI: 0.733-0.887), with a cutoff value of 18.35% for predicting poor outcomes, yielding a sensitivity of 67.57% and specificity of 92.16%.
Conclusion
For patients with spontaneous basal ganglia hemorrhage undergoing non-surgical treatment, admission SBP and the IHV/CSFV ratio were independently associated with poor prognosis in this cohort. As an exploratory composite imaging marker, IHV/CSFV was associated with outcome in our cohort and may represent a non-invasive, exploratory candidate marker for risk stratification that warrants further validation. Additionally, CSFV was significantly correlated with MTDTV, suggesting that CSFV could be effectively estimated through MTDTV measurements, a finding that could aid clinical decision-making.
Keywords: Spontaneous intracerebral hemorrhage, Basal ganglia, Cerebrospinal fluid volume, Hematoma volume, Maximum transverse diameter of the third ventricle
Introduction
Spontaneous intracerebral hemorrhage (ICH) refers to bleeding within the brain caused by non-traumatic factors. Primary causes include hypertension and cerebral amyloid angiopathy, while secondary factors involve intracranial aneurysms, cerebral arteriovenous malformations, and amyloidosis. ICH accounts for approximately 10–15% of all strokes, with most survivors left with varying degrees of neurological impairment [1].
Numerous studies have highlighted intracranial pressure (ICP) as a key factor influencing patient outcomes [2, 3]. According to the Monro-Kellie doctrine, ICP depends on the pressure-volume relationship among intracranial cerebrospinal fluid volume (CSFV), brain tissue, and blood, which can compensate for volume changes within certain limits [4]. However, when intracranial volume exceeds compensatory capacity, ICP increases exponentially [5–7], potentially leading to clinical manifestations such as headache, vomiting, and, in severe cases, brain herniation, respiratory arrest, or death.
While brain tissue occupies the largest volume within the cranial cavity, it cannot compress rapidly to adjust ICP. Instead, cerebrospinal fluid (CSF) plays a primary compensatory role. During ICP elevation, CSF is displaced from the cranial cavity into the subarachnoid space, with secretion reduced and absorption increased [8, 9]. This mechanism effectively lowers ICP. Thus, the volume of CSF is directly related to tolerance for elevated ICP and significantly impacts the outcomes of ICH patients.
CSF is generated in the ventricles and circulates into the subarachnoid space of the brain and spinal cord through the lateral and median apertures of the fourth ventricle [10]. Due to individual anatomical variations, intracranial CSFV differs widely between individuals, and precise measurement has historically been challenging. Early methods relied on invasive techniques, such as cadaver studies and pneumoencephalography [11], but advances in neuroimaging now allow non-invasive and accurate measurement through imaging modalities [12–14].
Emerging evidence has suggested that CSFV has predictive value for disease progression and therapeutic outcomes. For instance, the ratio of ischemic lesion volume to CSFV has been shown to accurately predict malignant middle cerebral artery infarction [15]. Building on this, our study employs CT imaging to measure CSFV and investigate the prognostic value of the ratio of intracerebral hematoma volume (IHV) to CSFV (IHV/CSFV) in non-surgically treated patients with spontaneous basal ganglia hemorrhage. This research aims to provide clinicians with evidence-based tools for decision-making and early identification of patients at risk for poor outcomes.
Materials and methods
Patients
With the approval of the Ethics Committee of our hospital, this study retrospectively enrolled non-surgically treated patients with spontaneous basal ganglia hemorrhage admitted to the Affiliated Hospital of Guizhou Medical University between January 2017 and October 2020. Patient data were retrieved from the hospital’s Health Information System (HIS).
Inclusion criteria
Patients with spontaneous basal ganglia hemorrhage who underwent head CT scans on the Siemens Somaris/5 syngo CT 2007 S at admission and did not receive surgical intervention.
Age range: 10–80 years.
Head CT performed within 24 h of hemorrhage onset.
Cases of initial bleeding with no subsequent rebleeding.
Exclusion criteria
Patients with hemorrhages secondary to other conditions, including aneurysms, intracranial tumors, or trauma.
Patients who underwent surgical interventions during hospitalization.
Patients diagnosed with hydrocephalus.
Patients presenting with intraventricular hemorrhage.
Patients on antithrombotic medication.
Patients with severe pre-existing physical or mental illnesses, or significant disabilities prior to the hemorrhage.
Pregnant patients. (Fig. 1.)
Fig. 1.
Study flowchart
Data collection
General patient information
Patient medical records were accessed via the HIS system. Following previous literature, factors potentially influencing patient outcomes were identified and collected. These included gender, age, the Glasgow Coma Scale (GCS) [16] score at admission (categorized into two groups: 0–12 and 13–15 points), the National Institute of Health Stroke Scale (NIHSS) [17] score at admission (categorized into two groups: 0–15 and 16–42 points), and blood pressure at admission. Three-month follow-up evaluations were conducted through telephone interviews or outpatient visits to obtain the patients’ modified Rankin Scale (mRS) [18] scores.
Cranial CT imaging
Cranial CT data were collected for all subjects. The CT scans were performed using a CT machine calibrated with the baseline aligned to the Orbitomeatal Line (OML). The imaging parameters included a scan field of 24.0 mm × 24.0 mm, a resolution of 512 × 512 pixels, and a slice thickness of 3 mm.
Measurement techniques
CSFV
For patients meeting the inclusion criteria, cranial CT image data were utilized to reconstruct 3-mm-thick slices with 3-mm intervals on a workstation for CSFV measurement. The Volume software integrated into the Siemens CT system was employed to calculate CSFV. Using this software, regions of interest were manually outlined across all slices, setting CT attenuation values to 0–20 Hounsfield units to isolate CSF. The software subsequently quantified CSFV in the subarachnoid space, ventricles, and various brain cisterns, collectively representing the total CSFV (Fig. 2).
Fig. 2.
Determination of cerebrospinal fluid volume (a) Circle selection of survey area and display of threshold area; b The result of the calculation of the regional content product by Volume
IHV and the maximum transverse diameter of the third ventricle(MTDTV)
The hematoma volume was measured using the ABC/2 formula, as proposed by Tada. Building on our previous studies [19, 20], which demonstrated a strong correlation between MTDTV and CSFV, this study further investigated the relationship between MTDTV, patient prognosis, and CSFV. The measurement process involved a radiologist and a neurosurgeon. By visual assessment, the imaging level with the largest MTDTV was identified. Using the distance measurement tool integrated into the workstation’s imaging software, the MTDTV was quantified (Fig. 3). To ensure reliability, measurements were conducted twice, and the mean value was recorded.
Fig. 3.
Measurement of the maximum transverse diameter of the third ventricle
Statistical analysis
Statistical analyses were conducted using SPSS version 26. The patient characteristics analyzed included gender, age, GCS score at admission, NIHSS score at admission, blood pressure at admission, IHV, CSFV, MTDTV, IHV/CSFV, IHV/MTDTV, and prognosis. Continuous data following a normal distribution were expressed as mean ± standard deviation and compared using independent sample t-tests. Non-normally distributed continuous data were presented as median and interquartile range [M (P25, P75)] and analyzed with the Mann-Whitney U test. Categorical variables were reported as frequencies and compared using the chi-square test.To identify significant influencing factors, logistic regression analysis was performed, providing odds ratios (ORs) with 95% confidence intervals (CIs). To address potential multicollinearity among closely related imaging metrics (IHV, CSFV, MTDTV, and IHV/CSFV), we constructed three separate hierarchical multivariable logistic regression models (Models 1, 2, and 3) rather than entering them simultaneously. Model 1 included IHV, Model 2 included CSFV, and Model 3 included the composite IHV/CSFV ratio. This approach allowed us to evaluate the association of each imaging construct with outcome while minimizing the direct confounding effects of collinearity. If the IHV/CSFV demonstrated an association with patient prognosis, a receiver operating characteristic (ROC) curve was constructed to determine its critical value for predicting an mRS score ≥ 3. The area under the curve (AUC), sensitivity, specificity, and other performance metrics were calculated. Additionally, correlation analyses were conducted to evaluate the relationships between CSFV and age, as well as CSFV and the MTDTV. Statistical significance was defined as P < 0.05 for all tests.
Results
Baseline characteristics
A total of 125 patients were included in the current study, of whom 51 (40.8%) achieved a good prognosis (mRS score ≤ 2), while 74 (59.2%) had a poor prognosis (mRS score ≥ 3). The cohort consisted of 82 male patients (65.6%) and 43 female patients (34.4%). Upon admission, 50 patients (40%) had a GCS score below 13, whereas 75 patients (60%) had a score of 13 or higher. Additionally, 88 patients (70.4%) presented with a NIHSS score under 16, while 37 patients (29.6%) scored 16 or above. (Table 1).
Table 1.
General information of patients
| Features | Number | Percentage(%) |
|---|---|---|
| Prognosis | ||
| mRS ≤ 2 | 51 | 40.8 |
| mRS ≥ 3 | 74 | 59.2 |
| Gender | ||
| Male | 82 | 65.6 |
| Female | 43 | 34.4 |
| GCS | ||
| <13 | 50 | 40.0 |
| ≥ 13 | 75 | 60.0 |
| NIHSS | ||
| <16 | 88 | 70.4 |
| ≥ 16 | 37 | 29.6 |
Univariate analysis
A normality test was performed on variables including patient age, SBP at admission, DBP at admission, CSFV, IHV/CSFV, the MTDTV, and IHV/MTDTV. The results indicated that patient age and SBP at admission followed a normal distribution, whereas the other variables did not. Consequently, t-tests were applied to variables with normal distributions, and non-parametric tests were used for those without normal distributions to assess their association with patient prognosis. Additionally, chi-square tests were utilized to analyze categorical variables such as gender, GCS score at admission, NIHSS score at admission, and IHV in relation to prognosis (Table 2).
Table 2.
Univariate analysis of prognosis
| Features | mRS(<3) | mRS(≥3) | X2/t/Z | P value |
|---|---|---|---|---|
| Gender | ||||
| Male | 36(70.6) | 46(62.2) | 0.950 | 0.330 |
| Female | 15(29.4) | 28(37.8) | ||
| GCS | ||||
| <13 | 5(9.8) | 45(60.8) | 32.729 | <0.001 |
| ≥ 13 | 46(90.2) | 29(39.2) | ||
| NIHSS | ||||
| <16 | 48(94.1) | 40(54.1) | 23.256 | <0.001 |
| ≥ 16 | 3(5.9) | 34(45.9) | ||
| IHV | 6.90 (4.00, 9.70) | 15.65 (9.10, 22.15) | -5.218 | <0.001 |
| Age | 55.31 ± 11.190 | 57.84 ± 11.092 | -1.246 | 0.215 |
| Admission SBP | 151.55 ± 20.364 | 170.93 ± 27.690 | -4.265 | <0.001 |
| Admission DBP | 89(81,100) | 101(91, 111) | -3.403 | <0.001 |
| CSFV | 71.6(61.2, 96.4) | 55.75(41.05, 72.375) | -3.843 | <0.001 |
| IHV/CSFV | 9.93(5.28, 14.36) | 30.22(14.13, 44.06) | -5.875 | <0.001 |
| MTDTV | 0.58(0.46, 0.80) | 0.40(0.31,0.6025) | -3.490 | <0.001 |
| IHV/ MTDTV |
127.60 (120.30, 139.03) |
134.36 (119.80, 152.23) |
-1.140 | 0.254 |
The results demonstrated that at 3 months post-hemorrhage, statistically significant differences (P < 0.05) were observed between patients with good and poor prognosis in terms of GCS score, NIHSS score, SBP, DBP, IHV, CSFV, IHV/CSFV, and MTDTV upon admission. In contrast, no significant differences were found concerning gender, age, or IHV/MTDTV (P > 0.05).
Multivariate logistic regression analysis
Independent variables with statistical significance in the univariate analysis (admission GCS, NIHSS, SBP, DBP, IHV, CSFV, IHV/CSFV, and MTDTV) were incorporated into hierarchical multivariable logistic regression models to identify independent predictors of poor prognosis. As shown in Table 3, SBP remained a consistent predictor across all three models (Model 1: OR = 1.056, P = 0.002; Model 2: OR = 1.050, P = 0.003; Model 3: OR = 1.053, P = 0.004), indicating that higher admission SBP was independently associated with an increased risk of poor outcome.
Table 3.
Hierarchical multivariable logistic regression
| Variable | Model1 | Model2 | Model3 | |||
|---|---|---|---|---|---|---|
| P | 95%CI | P | 95%CI | P | 95%CI | |
| GCS(per1) | 0.197 | 0.658(0.348,1.244) | 0.124 | 0.657(0.385,1.121) | 0.249 | 0.656(0.320,1.343) |
| NIHSS(per1) | 0.589 | 1.046(0.889,1.230) | 0.221 | 1.090(0.949,1.252) | 0.618 | 1.046(0.877,1.247) |
| SBP, mmHg(per1) | 0.002 | 1.056(1.019,1.093) | 0.003 | 1.050(1.017,1.084) | 0.004 | 1.053(1.017,1.091) |
| DBP, mmHg(per1) | 0.094 | 0.961(0.917,1.007) | 0.158 | 0.968(0.926,1.013) | 0.120 | 0.963(0.919,1.010) |
| MTDTV, cm(per1) | 0.019 | 0.087(0.011,0.675) | 0.455 | 0.148(0.001,22.239) | 0.543 | 0.508(0.057,4.506) |
| IHV, mL(per1) | 0.011 | 1.105(1.023,1.193) | — | — | — | — |
| CSFV, mL(per1) | — | 0.872 | 0.997(0.958,1.037) | — | — | |
| IHV/CSFV(per0.1) | — | — | 0.002 | 2.099(1.298,3.395) | ||
| — | — | — | — | |||
| N(included) | 125 | 125 | 125 | |||
| AUC | 0.884 | 0.871 | 0.888 | |||
| AIC | 117.496 | 124.614 | 113.066 | |||
| BIC | 137.294 | 144.413 | 132.864 | |||
Cells show OR (95% CI); P value. Model 1 includes GCS, NIHSS, SBP, DBP, MTDTV and IHV. Model 2 includes GCS, NIHSS, SBP, DBP, MTDTV and CSFV. Model 3 includes GCS, NIHSS, SBP, DBP, MTDTV and IHV/CSFV. AUC, AIC, and BIC are shown at the bottom
In the imaging-augmented models, IHV was an independent predictor in Model 1 (OR = 1.105, P = 0.011), whereas CSFV was not significant in Model 2 (OR = 0.997, P = 0.872). In Model 3, the IHV/CSFV ratio demonstrated a strong and significant association with poor prognosis (per 0.1 increase: OR = 2.099, P = 0.002), suggesting that IHV/CSFV may capture prognostic information not fully reflected by IHV or CSFV alone. Overall model performance further supported this finding: Model 3 showed numerically improved discrimination and model fit (AUC = 0.888;AIC = 113.066 and BIC = 132.864), supporting the potential incremental value of IHV/CSFV in this cohort. (Table 3).
Receiver Operating Characteristic (ROC) curve analysis
To evaluate the predictive performance of IHV/CSFV for poor prognosis, ROC curve analysis was performed (Fig. 4). IHV/CSFV demonstrated good discriminatory ability, with an AUC of 0.810 (95% CI: 0.733–0.887, P < 0.001). Using the Youden index, the optimal cutoff for predicting poor prognosis was 18.48%, suggesting that patients with IHV/CSFV > 18.48% had a markedly increased likelihood of poor outcome. At this threshold, the sensitivity was 67.57% and the specificity was 92.16% (Fig. 4).
Fig. 4.
ROC curve of IHV/CSFV and ICH-GS
In the same ROC comparison, the ICH grading scale(ICH-GS) [21] showed moderate predictive performance (AUC = 0.755, 95% CI: 0.669–0.841, P < 0.001), indicating that IHV/CSFV showed a higher AUC than the ICH-GS in this cohort, suggesting better discrimination; however, direct comparative testing and external validation are warranted (Fig. 4).
Correlation between CSFV, age, and MTDTV
Spearman correlation analysis was conducted to explore the relationship between CSFV, age, and MTDTV (Table 4). Results revealed a strong positive correlation between CSFV and MTDTV (r = 0.926, P < 0.01), supporting the feasibility of using MTDTV as a surrogate for estimating CSFV. In contrast, the correlation between CSFV and age was weak (r = 0.313, P < 0.01).
Table 4.
Spearman analysis of CSFV, age and maximum transverse diameter of the third ventricle
| CSFV | Age | MTDTV | |
|---|---|---|---|
| CSFV | 1.000 | 0.313** | 0.926** |
| Age | 0.313** | 1.000 | 0.319** |
| MTDTV | 0.926** | 0.319** | 1.000 |
Linear regression analysis
Scatter plots and residual analyses suggested a linear relationship between CSFV and MTDTV. Linear regression yielded the equation: CSFV = 117.7 × MTDTV + 6.948, providing a basis for estimating CSFV using MTDTV measurements.
Discussion
This study provides the first systematic evaluation of the prognostic value of the IHV/CSFV ratio in a well-defined cohort of patients with spontaneous basal ganglia hemorrhage managed non-surgically. By integrating hematoma burden with individual intracranial compensatory reserve, this exploratory composite imaging biomarker offers a potential new perspective for early risk stratification. Our findings demonstrate that the IHV/CSFV ratio is associated with three-month functional outcome: in our hierarchical regression analysis, the ratio showed a notable effect size and a numerically improved model fit (lowest AIC/BIC), suggesting its potential, hypothesis-generating utility in clinical prognostic assessment.
The IHV/CSFV ratio is grounded in a sound pathophysiological rationale. According to the Monro-Kellie doctrine, the intracranial compartment is of fixed volume; thus, an enlarging space-occupying lesion such as a hematoma must be offset by a reduction in either CSF or cerebral blood volume to maintain stable ICP [6]. CSFV, which reflects the brain’s compensatory capacity, varies considerably among individuals.While previous studies have largely focused on the mass effect of hematoma and the pathological changes it induces [22–28], the IHV/CSFV ratio—which normalizes hematoma volume to individual compensatory reserve—may provide a more comprehensive assessment of mass effect and its pathophysiological consequences than hematoma volume alone.Our hierarchical regression analysis (Table 3) lends empirical support to this hypothesis. When IHV and CSFV were entered as separate predictors (Models 1 and 2), CSFV did not reach statistical significance, and IHV showed only a modest predictive effect. However, when these two factors were combined into the IHV/CSFV ratio (Model 3), it emerged as a significant predictor (per 0.1 increase: OR = 2.099, P = 0.002), with the model demonstrating superior fit (lowest AIC/BIC). This pattern suggests that the prognostic information captured by the IHV/CSFV ratio is not merely additive but likely reflects an interaction between hematoma size and intracranial compensatory reserve—a higher ratio indicates a greater hematoma burden relative to the individual’s compensatory capacity, portending a disproportionately higher risk of critical ICP elevation and subsequent poor outcome.
Consistent with a large body of evidence linking elevated blood pressure to hematoma expansion and neurological deterioration [29–31], we confirmed that admission systolic blood pressure (SBP) is an independent predictor of poor prognosis. This finding reinforces the critical importance of strict early blood pressure control in this patient population, as recommended by current guidelines [32].
From a clinical translation perspective, the IHV/CSFV ratio may serve as a potential.
simple and effective, it does not account for individual variations in intracranial anatomy. In our cohort, the IHV/CSFV ratio demonstrated promising discriminatory ability in this cohort(AUC = 0.810), which was numerically higher than that of the ICH-GS (AUC = 0.755) (Fig. 4). This exploratory finding suggests that the ratio might capture additional prognostic information. However, we must emphasize that this does not demonstrate superior or incremental predictive value. Whether the IHV/CSFV ratio provides meaningful incremental prognostic information beyond well-established clinical scores such as the ICH-GS or ICH score requires rigorous testing in large, prospective, external validation cohorts. Future research should focus on whether incorporating this ratio into established models such as the ICH score can meaningfully improve predictive accuracy, thereby enabling more precise individualized prognostic assessment and treatment decision-making.
To facilitate potential clinical adoption, we explored the relationship between CSFV and a simple linear metric, the maximum transverse diameter of the third ventricle (MTDTV). The strong correlation observed (r = 0.926) and the derived linear regression equation (CSFV = 117.7 × MTDTV + 6.948) suggest that MTDTV, which can be rapidly and easily measured on routine non-contrast CT scans without specialized software, could serve as an accessible surrogate estimate for CSFV. While MTDTV cannot fully replace volumetric CSFV assessment, its potential use as a bedside screening tool may lower the technical barrier to applying this composite marker in routine clinical practice, particularly in resource-limited settings. Nevertheless, the clinical utility of this simplified approach requires independent prospective validation.
Limitations
Several limitations of this study warrant careful consideration when interpreting the results. First, as a single-center retrospective study with a relatively modest sample size (n = 125), the findings should be considered exploratory and require validation in larger, prospective, multicenter cohorts. Second, our stringent inclusion and exclusion criteria (excluding patients with intraventricular hemorrhage, hydrocephalus, or surgical intervention), while ensuring cohort homogeneity, limit the generalizability of our findings to the broader ICH population, particularly those with more severe hemorrhages. Third, regarding the measurement of hematoma volume, this study opted for the ABC/2 formula. This pragmatic methodological choice was based on clinical feasibility: the ABC/2 formula can be applied rapidly at the bedside without the need for complex post-processing, making it suitable for acute intracerebral hemorrhage diagnosis and treatment scenarios. However, it must be acknowledged that this method is less precise than volumetric segmentation and may have introduced measurement error. Ideally, both intracranial hematoma volume and cranial cavity volume should be quantified using automated or semi-automated segmentation tools to enhance accuracy and reproducibility. Fourth, we did not formally assess intra- or inter-observer reliability for CSFV and MTDTV measurements, representing a potential source of bias. Fifth, the inclusion of closely related variables (IHV, CSFV, MTDTV, and IHV/CSFV) in our models raises concerns about potential multicollinearity. While our hierarchical modeling approach, which entered IHV, CSFV, and IHV/CSFV into separate models, was specifically designed to mitigate this issue, it does not fully eliminate it. The observed strong performance of the IHV/CSFV ratio might still be influenced by its mathematical relationship with its components. Future studies with larger sample sizes should conduct formal collinearity diagnostics (e.g., variance inflation factors) and consider more robust techniques, such as ridge regression or bootstrapping, to further validate the independent contribution of the ratio. Sixth, dichotomization of the modified Rankin Scale, while clinically convenient, may obscure the full spectrum of disability. Future studies are encouraged to employ ordinal or shift analysis of the mRS to capture a more nuanced assessment of the marker’s prognostic performance across the full range of disability outcomes. Finally, the optimal cutoff value for IHV/CSFV identified in this study (18.48%) is data-driven and requires prospective validation in independent external populations before being considered for clinical decision-making.
Conclusion
In conclusion, this exploratory study provides preliminary evidence that the IHV/CSFV ratio—an imaging biomarker integrating hematoma burden with intracranial compensatory reserve—is significantly associated with three-month functional outcome in patients with spontaneous basal ganglia hemorrhage undergoing non-surgical treatment. This novel metric holds promise as a potential reference indicator for early clinical risk stratification. Its strong correlation with the easily measurable MTDTV further facilitates its potential translation into clinical practice. It must be emphasized, however, that these findings are hypothesis-generating. The incremental prognostic value of the IHV/CSFV ratio beyond established scoring systems remains to be validated. Large-scale, multicenter, prospective studies are warranted to validate the prognostic value of the IHV/CSFV ratio, assess its incremental contribution, and ultimately determine its role in guiding individualized treatment decisions for this complex patient population.
Acknowledgements
Not applicable.
Authors' contributions
Zhangwei Yan, Xu Xu, Hua Yang and Xin Xiang designed the project. Xingwang Zhou, Xiaoyu Wang and Tao Sun collected and analyzed the data. Zhangwei Yan drafted the manuscript. Hua Yang, Xin Xiang and Junshuan Cui revised the manuscript. All authors approved the final version of the manuscript.
Funding
We express our heartfelt thanks to the dedicated staff of the Department of Neurosurgery at The Affiliated Hospital of Guizhou Medical University for their invaluable support. Our research was generously funded by the National Natural Science Foundation of China (Grant No. 82560249) and the 2023 Central Government Subsidy Fund allocated for Enhancing Medical Services and Security Capabilities, specifically targeting the development of Medical and Health Institutions (National Key Clinical Specialist - Neurosurgery, Grant No. 2023-95). Additionally, we are immensely grateful to the patients who kindly consented to provide pathological specimens for our study. Their contributions were absolutely essential to the realization of this research.
Data availability
The datasets used and/or analyzed during the present study are available from the authors on reasonable request. Code used throughout this study is available upon reasonable request from the corresponding authors.
Declarations
Ethics approval and consent to participate
This study has been approved by the Ethics Committee of the Affiliated Hospital of Guizhou Medical University.All procedures performed in this study involving human participants were in accordance with the Declaration of Helsinki (as revised in 2013) and the Ethical Guidelines for Medical Research Involving Human Subjects issued by the National Health Commission of the People's Republic of China.
All patients included in this study (or their legal representatives, if the patient was unable to provide consent due to neurological impairment caused by acute basal ganglia hemorrhage) signed written informed consent forms prior to the use of their clinical data and imaging materials. For patients with impaired consciousness at admission, informed consent was obtained from their legal next of kin, and additional written consent was obtained from the patients themselves once they regained the capacity to communicate and understand the study purpose.
Consent for publication
Not applicable.
Competing interests
The authors declare no competing interests.
Footnotes
Publisher’s note
Springer Nature remains neutral with regard to jurisdictional claims in published maps and institutional affiliations.
Contributor Information
Hua Yang, Email: yhmed@163.com.
Xin Xiang, Email: xiangxin828@163.com.
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Associated Data
This section collects any data citations, data availability statements, or supplementary materials included in this article.
Data Availability Statement
The datasets used and/or analyzed during the present study are available from the authors on reasonable request. Code used throughout this study is available upon reasonable request from the corresponding authors.